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On the Identification of Production Functions: How
Heterogeneous is Productivity?
Amit Gandhi, Salvador Navarro, David Rivers∗
May 30, 2016
Abstract
We show that the “proxy variable” approaches for estimating production functions generi-
cally suffer from a fundamental identification problem in the presence of a flexible input that
satisfies the “proxy variable” assumptions for invertibility, such as intermediate inputs. We
provide a formal proof of nonparametric non-identification, and illustrate that the source of
under-identification is the flexible input elasticity. Using a transformation of the firm’s first
order condition, we develop a new nonparametric identification strategy that addresses this
problem, as well as a simple corresponding estimator for the production function. We show
that the alternative of approximating the effects of intermediate inputs using a value-added
production function cannot generally be used to identify features of interest from the gross
output production function. Applying our approach to plant-level data from Colombia and
Chile, we find that a gross output production function implies fundamentally different patterns
of productivity heterogeneity than a value-added specification.
∗We would like to thank Dan Ackerberg, Richard Blundell, Juan Esteban Carranza, Allan Collard-Wexler, UlrichDoraszelski, Steven Durlauf, Jeremy Fox, Silvia Goncalves, Phil Haile, Joel Horowitz, Jean-Francois Houde, Au-reo de Paula, Amil Petrin, Mark Roberts, Nicolas Roys, Chad Syverson, Chris Taber, Quang Vuong, and especiallyTim Conley for helpful discussions. This paper has also benefited from detailed comments by the editor and threeanonymous referees. We would also like to thank Amil Petrin and David Greenstreet for helping us to obtain theColombian and Chilean data respectively. Navarro and Rivers acknowledge support from the Social Sciences andHumanities Research Council of Canada. This paper previously circulated under the name “Identification of Produc-tion Functions using Restrictions from Economic Theory.” First draft: May 2006. Amit Gandhi is at the Universityof Wisconsin-Madison, E-mail: [email protected]. Salvador Navarro is at the University of Western Ontario,E-mail: [email protected]. David Rivers is at the University of Western Ontario, E-mail: [email protected].
1
1 Introduction
The identification and estimation of production functions using data on firm inputs and output
is among the oldest empirical problems in economics. A key challenge for identification arises
because firms optimally choose their inputs as a function of their productivity, but productivity is
unobserved by the econometrician in the data. This gives rise to a classic simultaneity problem that
was first articulated by Marschak and Andrews (1944) and has come to be known in the production
function literature as “transmission bias”. In their influential review of the state of the literature
nearly 50 years later, Griliches and Mairesse (1998) (henceforth GM) concluded that the “search
for identification” for the production function remained a fundamentally open problem.1 Resolving
this identification problem is critical to measuring productivity with plant level production data,
which has become increasingly available for many countries, and which motivates a variety of
industry equilibrium models based on patterns of productivity heterogeneity found in this data.2
In this paper we examine the identification foundations of the “proxy variable” approach for
estimating production functions pioneered by Olley and Pakes (1996) and further developed by
Levinsohn and Petrin (2003)/Ackerberg, Caves, and Frazer (2015)/Wooldridge (2009) (henceforth
OP, LP, ACF and Wooldridge, respectively). Despite the immense popularity of the method in
the applied productivity literature, a fundamental question has remained unexplored: does the
proxy variable technique solve the identification problem of transmission bias? That is, is the
production function in fact identified from panel data on inputs and output under the structural
assumptions that the proxy variable technique places on the data? To date there has been no formal
demonstration of identification in this literature. On the other hand, there have been various red
flags concerning the identification foundations for this class of estimators (see Bond and Söderbom,
2005 and ACF). However, no clear conclusion regarding non-identification of the model as a whole
1In particular, the standard econometric solutions to correct the transmission bias, i.e., using firm fixed effects orinstrumental variables, have proven to be both theoretically problematic and unsatisfactory in practice (see e.g., GMand Ackerberg et al., 2007 for a review and Section 7 of this paper for a discussion).
2Among these patterns are the general understanding that even narrowly defined industries exhibit “massive” unex-plained productivity dispersion (Dhrymes, 1991; Bartelsman and Doms, 2000; Syverson, 2004; Collard-Wexler, 2010;Fox and Smeets, 2011), and that productivity is closely related to other dimensions of firm-level heterogeneity, such asimporting (Kasahara and Rodrigue, 2008), exporting (Bernard and Jensen, 1995, Bernard and Jensen, 1999, Bernardet al., 2003), wages (Baily, Hulten, and Campbell, 1992), etc. See Syverson (2011) for a review of this literature.
2
has been reached either. Without an answer to this basic question, it is unclear how to interpret the
vast body of empirical results regarding patterns of productivity which rely on the proxy approach
for the measurement of productivity.
Our first main result is a negative one. We show that the model structure underlying the proxy
variable technique does not identify the production function (and hence productivity) in the pres-
ence of a flexible input that satisfies the conditions for invertibility invoked in this literature. We
establish this result under the LP/Wooldridge approach that uses intermediate inputs as a flexi-
ble input that satisfies the proxy variable assumption of being monotone (and hence invertible) in
productivity. Under this structure we show how to construct a continuum of observationally equiv-
alent production functions that cannot be distinguished from the true production function in the
data. We also show that similar non-identification problems arise under the original OP strategy of
using investment as the proxy variable.3
Our second contribution is to show that the exact source of this non-identification is the flexible
input elasticity for inputs satisfying the proxy variable assumption. The flexible input elasticity
defines a partial differential equation on the production function. As we show, if this elasticity
were known, then it could be integrated up to nonparametrically identify the part of the production
function that depends on the flexible input. We then show that the proxy variable structure is
sufficient to nonparametrically identify the remainder of the production function. These results
taken together formalize the empirical content of the proxy variable structure that is widely used
in applied work - the model is identified up to the flexible input elasticity, but fails to identify the
flexible input elasticity itself.
Our third contribution is that we present a new empirical strategy that nonparametrically iden-
tifies the flexible input elasticity, and hence solves for the missing source of identification for the
production function within the proxy variable structure. The key to our approach is that we build
on a key economic assumption of the proxy variable structure - that the firm optimally chooses
intermediate inputs in response to its realized productivity. Whereas the proxy variable literature
3We do not emphasize OP as the leading case because the applied literature has largely adopted the use of interme-diate inputs as the proxy variable due to the prevalence of zeroes in investment, which was the original motivation forLP. Nevertheless we show that resorting to OP does not solve the non-identification problem we pose for LP.
3
has used this assumption to “invert” and replace for productivity using intermediate inputs, we go
further and exploit the nonparametric structural link between the production function and the firm’s
first order condition for intermediate inputs. This link allows for a nonparametric regression of the
intermediate input’s revenue share on all inputs (labor, capital, and intermediate inputs) to identify
the flexible input elasticity. This is a nonparametric analogue of the familiar parametric insight that
revenue shares directly identify the intermediate input coefficient in a Cobb-Douglas setting (e.g.,
Klein, 1953 and Solow, 1957). Our key innovation is that we show that the information in the first
order condition can be used in a completely nonparametric way, i.e., without making functional
form assumptions on the production function. Our results thus show how this “share regression”
can be combined with the remaining content of the proxy variable structure to nonparametrically
identify the production function as a whole.
This identification strategy - regressing revenue shares on inputs to identify the flexible input
elasticity, solving the partial differential equation, and integrating this into the proxy variable struc-
ture to identify the remainder of the production function - gives rise to a natural two-step estimator
in which a flexible parametric approximation to different components of the production function is
estimated in each stage. We present a computationally straightforward implementation of this es-
timator and show that the properties of the estimator are equivalent to a standard GMM estimator,
which gives us a straightforward approach to inference with the computed parameter estimates.
We further relate our solution to the common empirical practice in the literature of estimating
value-added production functions that subtract out a flexible input from the model (typically inter-
mediate inputs). While value-added specifications may be of direct interest themselves, a common
justification for value-added is that they derive directly from an underlying gross output technol-
ogy. To the extent that one is interested in features of the gross output production function, value
added production functions may appear immune from the non-identification problem we raise,
as they explicitly exclude intermediate inputs. We show that, unless the production function is a
very specific version of Leontief in value added and intermediate inputs, then value added cannot
be used to identify features of interest (including productivity) from the underlying gross output
production.
4
Finally, we apply our identification strategy to plant-level data from Colombia and Chile to
study the underlying patterns of productivity under gross output compared to value-added specifi-
cations. We find that productivity differences become orders of magnitude smaller and sometimes
even change sign when we analyze the data via gross output rather than value added. For example,
the standard 90/10 productivity ratio taken among all manufacturing firms in Chile is roughly 9
under value added (meaning that the 90th percentile firm is 9 times more productive than the 10th
percentile firm), whereas under our gross output estimates this ratio falls to 2. Moreover, these dis-
persion ratios exhibit a remarkable degree of stability across industries and across the two countries
when measured via gross output, but exhibit much larger cross-industry and cross-country variance
when measured via value added. We further show that, as compared to gross output, value added
estimates generate economically significant differences in the productivity premium of firms that
export, firms that import, firms that advertise, and higher wage firms.
In contrast to the view expressed in Syverson (2011), that empirical findings related to produc-
tivity are quite robust to measurement choices, our findings illustrate the empirical importance of
the distinction between gross output and value-added estimates of productivity. Our results high-
light the empirical relevance of our identification strategy for gross output production functions.
The results suggest that the distinction between gross output and value added is at least as im-
portant, if not more so, than the transmission bias that has been the main focus of the production
function estimation literature to date.
The rest of the paper is organized as follows. In Section 2 we describe the model and provide
a nonparametric characterization of transmission bias. In Section 3 we formally prove that the
production function is nonparametrically non-identified under the proxy variable approach. Sec-
tion 4 shows that the source of the non-identification is the flexible input elasticity. In Section 5
we present our nonparametric identification strategy. Section 6 describes our estimation strategy.
Section 7 compares our approach to the related literature. Section 8 discusses the use of value
added. In Section 9 we describe the Colombian and Chilean data and show the results compar-
ing gross output to value added for productivity measurement. In particular, we show evidence
of large differences in unobserved productivity heterogeneity suggested by value added relative to
5
gross output. Section 10 concludes with an example of the policy relevance of our results.
2 The Model and Identification Problem
We first describe the economic model of production underlying the “proxy variable” approach
to estimating production functions (OP/LP/ACF/Wooldridge), which has become a widely-used
approach to estimating production functions and productivity in applied work. We then define
the identification problem associated with the model and data. Our first main result in this pa-
per demonstrates that, despite the ubiquity of these approaches in empirical work, the production
function and productivity are not identified by the restrictions imposed by these methods.4,5
2.1 Data and Definitions
We observe a panel consisting of firms j = 1, . . . , J over periods t = 1, . . . , T .6 A generic firm’s
output, labor, capital, and intermediate inputs will be denoted by (Yt, Kt, Lt,Mt) respectively,
and their log values will be denoted in lowercase by (yt, kt, lt,mt). Firms are sampled from an
underlying population and the asymptotic dimension of the data is to let the number of firms
J → ∞ for a fixed T , i.e., the data takes a short panel form. The data directly identifies the
joint distribution of the history of inputs and output of a firm, i.e., the data identifies the joint the
distribution of the collection of random variables: {(yt, kt, lt,mt)}Tt=1 .
We let It denote the information set of the firm in period t. The information set It consists of all
information the firm can use to solve its period t decision problem, which potentially includes its
input choices. By definition (and irrespective of the economics details underlying input decisions)
we have that the capital, labor, and intermediate input choices in each period can be expressed as
4As we discuss in Sections 3 and 8, Ackerberg, Caves, and Frazer (2015) avoid the issues we discuss below bycarefully considering data-generating-processes under which their procedure can be employed for restricted profit /Leontief specifications of the production function.
5The identification problem we isolate also applies to the dynamic panel approach to production function estimationfollowing Arellano and Bond (1991); Blundell and Bond (1998, 2000). We draw a comparison to the dynamic panelliterature in Section 7.3.
6Throughout this section we assume a balanced panel for notational simplicity. We also omit the firm subscript j,except when the context requires it.
6
functions of the information set It, i.e.,
kt = K (It) ; lt = L (It) ; mt = M (It) (1)
Let xt ∈ {kt, lt,mt} denote a generic input. If an input xt is such that xt ∈ It, i.e., the period
t amount of the input employed in the period is in the firm’s information set for that period, then
we say the input is predetermined in period t. Thus a predetermined input is a function of the
information set of a prior period,
xt = X (It−1) ∈ It
If an input is not predetermined, and thus xt /∈ It, then we say the input is variable in period t.
If an input is variable and∂
∂xt−τX (It) 6= 0
for τ > 0, i.e., lagged values of the input affect the optimal period t choice of the input, then we
say the input is dynamic. Finally, if a variable input is not dynamic, then we say it is flexible.
2.2 The Production Function and Productivity
We assume that the relationship between output and inputs is determined by an underlying produc-
tion function F and a Hicks neutral productivity shock νt.
Assumption 1. The relationship between output and the inputs takes the form
Yt = F (kt, lt,mt) eνt ⇐⇒
yt = f (kt, lt,mt) + νt (2)
The Hick’s neutral productivity shock νt is decomposed as νt = ωt+εt. The distinction between
ωt and εt is that ωt is known to the firm before making its period t decisions, whereas εt is an ex-
post productivity shock realized only after the period decisions are made. The stochastic behavior
of both of these components is explained next.
7
Assumption 2. ωt ∈ It is known to the firm at the time of making its period t decisions, whereas
εt /∈ It is not. Furthermore ωt is Markovian so that its distribution can be written as Pω (ωt | It−1) =
Pω (ωt | ωt−1). The function h (ωt−1) = E [ωt | ωt−1] is continuous. The shock εt on the other hand
is independent of the within period variation in information sets, Pε (εt | It) = Pε (εt).
Given that ωt ∈ It, but εt is completely unanticipated on the basis of It, we will refer to ωt as
persistent productivity, εt as ex-post productivity, and νt = ωt + εt as total productivity. Observe
that we can express ωt = h(ωt−1) + ηt, where ηt satisfies E [ηt | It−1] = 0. ηt can be interpreted as
the, unanticipated at period t− 1, “innovation” to the firm’s persistent productivity ωt in period t.7
Without loss of generality, we can normalize E [εt | It] = E [εt] = 0, which is in units of log
output. However, the expectation of the ex-post shock, in units of the level of output, becomes a
free parameter which we denote as E ≡ E [eεt | It] = E [eεt ].8 Note that the general form of input
demand (1) implies E [εt | It, kt, lt,mt] = E [εt | It] = 0 and hence E [εt | kt, lt,mt] = 0 by the
law of iterated expectations.
Finally, we have the scalar invertibility assumption that allows an input to be used to proxy for
productivity. Levinsohn and Petrin (2003) proposed using the structure of a flexible input demand
as the basis for such a proxy variable. For simplicity, we focus on the case of a single flexible input
in the model, namely intermediate inputs mt, and treat capital kt and labor lt as predetermined in
the model. The non-identification problem we demonstrate can be easily adapted to the case where
lt is also flexible.9
Assumption 3. The scalar unobservability assumption of LP/Wooldridge places the following as-
7It is straightforward to allow the distribution of Pω (ωt | It−1) to depend upon other elements of It−1, such asfirm export or import status, R&D, etc. In these cases ωt becomes a controlled Markov process from the firm’s pointof view. See Kasahara and Rodrigue (2008) and Doraszelski and Jaumandreu (2013) for examples.
8See Goldberger (1968) for an early discussion of the implicit reinterpretation of results that arises from ignoringE (i.e., setting E≡ E [eεt ] = 1 while simultaneously setting E [εt] = 0) in the context of Cobb-Douglas productionfunctions.
9We focus on the case of a single flexible input because allowing lt to be flexible, in addition to intermediate inputsmt, is associated with a set of distinct problems related to the identification of the labor elasticity that were raised byACF. The key to avoiding the ACF critique is letting labor have sources of variation beyond ωt and yet preserve thescalar unobservability restriction on the intermediate input demand. By treating lt as predetermined we avoid the ACFcritique and can focus attention on the flexible input elasticity problem (which is the focus of our paper).
8
sumption on the flexible input demand
mt = M (It) = M (kt, lt, ωt) . (3)
The intermediate input demand M is assumed strictly monotone in ωt.10
Observe that a key implication of Assumption 3 is that M can be inverted in ωt, allowing
productivity to be expressed as a deterministic function of the inputs
ωt = M−1 (kt, lt,mt) .
We restrict our attention to the use of intermediate inputs as a proxy versus the original proxy
variable strategy of OP that uses investment. As LP argued, the fact that investment is often zero
in plant level data leads to practical challenges in using the OP approach, and as a result using
intermediate inputs as a proxy has become the preferred strategy in applied work. Investment as a
proxy raises similar identification challenges, which we discuss in Appendix A.
We could have generalized the model to allow the primitives to vary with time t, i.e., ft, Pt,ω,
and Pt,ε to all vary by time t. We do not use this more general form of the model in the analysis to
follow because the added notational burden distracts from the main ideas of the paper. However,
it is straightforward to generalize the analysis that follows to the time-varying case by simply
repeating the steps of our analysis separately for each time period t ∈ {2, . . . , T}.
2.3 Transmission Bias
Given the structure of the production function we can formally state the problem of transmission
bias in the nonparametric setting. Transmission bias classically refers to the bias of the OLS re-
gression of output on inputs as estimates of a Cobb-Douglas production parameter. In the nonpara-
metric setting we can see transmission bias more generally as the empirical problem of regressing
10This approach can be generalized to allow the input demand M to vary by time period t. All of our results can bereadily extended to this more general case at the cost of introducing additional notational complexity, which we avoidhere in order not to distract from the core conceptual issues we raise.
9
output yt on inputs (kt, lt,mt) which yields
E [yt | kt, lt,mt] = f (kt, lt,mt) + E [ωt | kt, lt,mt]
and hence the elasticity of the regression in the data with respect to an input xt ∈ {kt, lt,mt}
∂
∂xtE [yt | kt, lt,mt] =
∂
∂xtf (kt, lt,mt) +
∂
∂xtE [ωt | kt, lt,mt]
is a biased estimate of the true production elasticity ∂∂xtf (kt, lt,mt).
Under the proxy variable structure, transmission bias takes a very specific form. This can be
seen as follows:
E [yt | kt, lt,mt] = f (kt, lt,mt) + M−1 (kt, lt,mt) ≡ φ (kt, lt,mt) . (4)
Clearly no structural elasticities can be identified from this regression (the “first stage”), in partic-
ular the flexible input elasticity, ∂∂mt
f (kt, lt,mt). Instead, all the information from the first stage is
summarized by the identification of the random variable φt ≡ φ (kt, lt,mt), and as a consequence
the ex-post productivity shock εt = yt − E [yt | kt, lt,mt].
The question then becomes whether the part of φt that is due to f (kt, lt,mt) versus the part
due to ωt can be separately identified using the second stage restrictions of the model. This second
stage is formed by recognizing that
yt = f (kt, lt,mt) + ωt + εt
= f (kt, lt,mt) + h (φt−1 − f (kt−1, lt−1,mt−1)) + ηt + εt. (5)
The challenge in using this equation for identification is the presence of an endogenous variable
mt in the model that is correlated with ηt.
LP/Wooldridge propose to use instrumental variables for this endogeneity problem by exploit-
ing orthogonality restrictions implicit in the model with respect to ηt+εt. In particular Assumption
10
2 implies that for any transformation Γt = Γ (It−1) of the lagged period information set It−1 we
have the orthogonality E [ηt + εt | Γt] = 0. We focus on transformations that are observable by
the econometrician, in which case Γt will serve as the instrumental variables for the problem. The
full vector of potential instrumental variables given the data (as described in section 2.1) consists
of all lagged output/input values which, by construction, are transformations of It−1, as well as the
current values of the predetermined inputs kt and lt (which by assumption are a transformation of
It−1), i.e., Γt = (kt, lt, yt−1, kt−1, lt−1,mt−1, . . . , y0, k0, l0,m0).
3 Non-Identification
In this section we show that the proxy variable structure of Assumptions 1 - 3 does not suffice to
identify the production function. In Theorem 1, we first show that the application of instrumental
variables (via the orthogonality restriction E [ηt + εt | Γt] = 0) to the structural equation (5) is
insufficient to identify the production function f (and the Markovian process h). However, the
orthogonality restriction underlying the instrumental variables approach does not summarize the
full structure of the model. In Theorem 2, we show that, even if we treat the model as a simultane-
ous system of equations for the determination of the the endogenous variables (yt,mt) in (5), the
production function f cannot be identified.
Identification of the production function f by instrumental variables is based on projecting
output yt onto the exogenous variables Γt (see e.g., Newey and Powell, 2003). This generates a
restriction between (f, h) and the distribution of the data Gyt,mt|Γt that takes the form
E [yt | Γt] = E [f (kt, lt,mt) | Γt] + E [ωt | Γt]
= E [f (kt, lt,mt) | Γt] + h (φt−1 − f (kt−1, lt−1,mt−1)) , (6)
where recall that φt−1 ≡ φ (kt−1, lt−1,mt−1) is known from the first stage equation (4). The
structural primitives underlying equation (6) are given by (f, h). The true (f 0, h0) are identified if
no other(f , h
)among all possible alternatives also satisfy the functional restriction (6) given the
11
distribution of the observables Gyt,mt|Γt .11
We first establish the following useful Lemma. For notational simplicity, we define the random
variable ft ≡ f (kt, lt,mt).
Lemma 1. If (f, h) solve the functional restriction (6), then it must be the case that
E [φt − ft | Γt] = h (φt−1 − ft−1)
Proof. Observe that
E [yt | Γt] = E [E [yt | kt, lt,mt] | Γt]
= E [φt | Γt]
by construction of φt. From the definition of yt it follows that
E [φt | Γt] = E [ft | Γt] + h (φt−1 − ft−1) .
Re-arranging terms gives us the Lemma.
Theorem 1. Under the model defined by Assumptions 1 - 3, and given φt ≡ φ (kt, lt,mt) identified
from the first stage equation (4), there exists a continuum of alternative(f , h
)defined by
f ≡ (1− a) f 0 + aφt
h (x) ≡ (1− a)h0
(1
(1− a)x
)
for any a ∈ (0, 1), that satisfy the same functional restriction (6) as the true (f 0, h0).
Proof. The proof of the Theorem follows almost immediately from Lemma 1. Given the definition
11Less formally, the intuitive idea is that(f0, h0
)are the unique primitives that explain the reduced form E [yt | Γt]
given the model.
12
of(f , h
)we have
ft + h(φt−1 − ft−1
)=
f 0t + a
(φt − f 0
t
)+ h
((1− a)
(φt−1 − f 0
t−1
))=
f 0t + a
(φt − f 0
t
)+ (1− a)h0
(φt−1 − f 0
t−1
).
Now, take the conditional expectation of the above (with respect to Γt) and apply the Lemma
E[ft | Γt
]+ h
(φt−1 − ft−1
)=
E[f 0t | Γt
]+ ah0
(φt−1 − f 0
t−1
)+ (1− a)h0
(φt−1 − f 0
t−1
)=
E[f 0t | Γt
]+ h0
(φt−1 − f 0
t−1
).
Thus (f 0, h0) and(f , h
)satisfy the functional restriction and cannot be distinguished via instru-
mental variables.
The intuition for the identification failure established in Theorem 1 can be seen by looking at
equation (5) above. Notice that, by replacing for ωt in the intermediate input demand equation (3)
we get
mt = M(kt, lt, h
(M−1 (kt−1, lt−1,mt−1)
)+ ηt
).
This implies that the only source of variation left in mt after conditioning on
(kt, lt, kt−1, lt−1,mt−1) ∈ Γt (which are used as instruments for themselves) is the unobservable ηt.
Therefore, despite the apparent abundance of instruments in Γt, all of the remaining elements in Γt
are orthogonal to this remaining source of variation, ηt, and hence have no power as instruments.
Theorem 1 calls into question existing applied work on productivity that employs the proxy
variable technique to recover productivity. A standard application in the literature will employ a
“flexible parametric approximation” fβ parametrized by a finite dimensional parameter β to the
production function f . The standard estimator applies the restrictions of the first stage and second
stage in a GMM formulation where the moments are defined by E [εtxt] = 0 for xt ∈ {kt, lt,mt}
13
and E [(ηt + εt)xt] for xt ∈ Γt. So long as the number of moment restrictions (determined by
the number of elements in Γt the estimator exploits) exceeds the dimensionality of β, then such
an approach would appear identified. However, our Theorem 1 shows that this simple process of
“counting moment equations” and unknown parameters is deceiving.
In so far as such estimators are consistent for β, it is only because of the parametric structure
being employed. Theorem 1 establishes that the production function is nonparametrically non-
identified by the moment restrictions underlying these techniques. However, the researcher will
typically have little basis for imposing parametric restrictions, and, if the parametric restrictions are
not correct, this can generate misleading inferences about the production function and productivity
(see Manski, 2003; Roehrig, 1988; and Matzkin, 2007 for more detail). Furthermore, as we show
in Appendix B, for the case of the commonly-employed Cobb-Douglas parametric form, even
imposing structural parametric assumptions is not necessarily sufficient to solve the identification
problem.
The result in Theorem 1 is a useful benchmark, as it relates directly to the approach used in the
proxy variable literature. However, this instrumental variables approach does not necessarily ex-
haust the sources of identification inherent in the proxy variable structure. First, since instrumental
variables is based only on conditional expectations, it does not employ the entire distribution of the
data (yt,mt,Γt). Second, it does not directly account for the fact that Assumption 3 also imposes
restrictions (scalar unobservability and monotonicity) on the determination of the endogenous vari-
able mt via M (·). Therefore, the proxy variable structure imposes restrictions on a simultaneous
system of equations because, in addition to the model for output, yt, via the production function,
there is a model for the proxy variable, in this case intermediate inputs, mt.
We now extend our non-identification result to show that the complete structure Θ = (f, h,M)
cannot be identified from the full joint distribution Gyt,mt|Γt of the data. For a structure Θ, let
εΘt = yt − f (kt, lt,mt)−M−1 (kt, lt,mt) ,
14
and
ηΘt = M−1 (kt, lt,mt)− h
(M−1 (kt−1, lt−1,mt−1)
).
In order to relate the structure Θ to the joint distribution of the data Gyt,mt|Γt through the model, a
joint distribution of the GηΘt ,ε
Θt |Γt needs to be specified. Let EG (·) denote the expectation operator
taken with respect to distribution G. We say that a structure Θ rationalizes the data if: (i) there
exists a joint distribution GηΘt ,ε
Θt |Γt = GηΘ
t |Γt × GεΘtthat generates the joint distribution Gyt,mt|Γt;
(ii) satisfies the first stage moment restriction EGεΘt
[εΘt | kt, lt,mt
]= 0; (iii) satisfies the IV or-
thogonality restriction EGηΘt ,ε
Θt |Γt
[ηΘt + εΘ
t | Γt]
= 0; and (iv) satisfies Assumption 3 (i.e., scalar
unobservability and monotonicity of M). Following Matzkin (2007), we say that, if there exists an
alternative structure Θ 6= Θ0 that rationalizes the data, then the structure Θ0 is not identified from
the joint distribution Gyt,mt|Γt of the data.
Theorem 2. Given the true structure Θ0 = (f 0, h0,M0) and Assumptions 1 - 3, there always exists
a continuum of alternative structures Θ 6= Θ0, defined by
f ≡ f 0 + a(M0)−1
h (x) ≡ (1− a)h0
(1
(1− a)x
)M−1 ≡ (1− a)
(M0)−1
for any a ∈ (0, 1), that exactly rationalize the data Gyt,mt|Γt .
Proof. Let x denote a particular value of the random variable x in its support. We first observe that,
for any hypothetical structure Θ = (f, h,M), there always exists a distribution GηΘt ,ε
Θt |Γt defined
by
GηΘt ,ε
Θt |Γt (ηt, εt | Γt) =
Gyt,mt|Γt
εt + f (kt, lt,M (kt, lt, h (M−1 (kt−1, lt−1,mt−1)) + ηt))
+M−1 (kt, lt,M (kt, lt, h (M−1 (kt−1, lt−1,mt−1)) + ηt))
,M (kt, lt, h (M−1 (kt−1, lt−1,mt−1)) + ηt)
∣∣∣∣∣∣∣∣∣Γt ,
15
that generates the conditional distribution of the data Gyt,mt|Γt through the model, hence (i) is
satisfied.
Second, since the true model rationalizes the data, it follows that EGεΘ
0t
[εΘ0
t | kt, lt,mt
]= 0.
The εΘt implied by our alternative structure is given by
εΘt = yt − f (kt, lt,mt)− M−1 (kt, lt,mt)
= yt − f 0 (kt, lt,mt)− a(M0)−1
(kt, lt,mt)− (1− a)(M0)−1
(kt, lt,mt)
= yt − f 0 (kt, lt,mt)−(M0)−1
(kt, lt,mt)
= εΘ0
t ,
so it trivially satisfies the moment restriction in (ii).
Third, it follows that
ηΘt + εΘ
t = yt − f (kt, lt,mt)− h(M−1 (kt−1, lt−1,mt−1)
)= yt − f 0 (kt, lt,mt)− a
(M0)−1
(kt, lt,mt)− (1− a)h0((
M0)−1
(kt−1, lt−1,mt−1))
= yt − f 0 (kt, lt,mt)− h0((
M0)−1
(kt−1, lt−1,mt−1))
︸ ︷︷ ︸ηΘ0t +εΘ
0t
−a
(M0)−1
(kt, lt,mt)− h0((
M0)−1
(kt−1, lt−1,mt−1))
︸ ︷︷ ︸ηΘ0t
= (1− a) ηΘ0
t + εΘ0
t .
Since εΘt = εΘ0
t , it immediately follows that EGηΘt ,ε
Θt |Γt
(εΘt | Γt
)= 0. It also follows that ηΘ
t =
16
(1− a) ηΘ0
t . By a simple change of variables we have that
EGηΘt ,ε
Θt |Γt
(ηΘt | Γt
)= EG
ηΘt |Γt
(ηΘt | Γt
)= EG
ηΘ0t |Γt
(ηΘt
(1− a)| Γt
)= EG
ηΘ0t |Γt
(ηΘ0
t | Γt)
= 0.
Hence, our alternative structure satisfies the moment restriction in (iii).
Finally we notice that, since (M0)−1 is invertible given Assumption 3,
(M)−1
≡ (1− a) (M0)−1
is therefore also invertible and hence satisfies Assumption 3 (i.e., (iv)) as well. Since both Θ and
Θ0 satisfy requirements (i)-(iv), i.e., both rationalize the data, we conclude that Θ0 is not identi-
fied.
While we have focused our discussion on gross output production functions, a similar non-
identification result generally arises in a value-added specification in which either capital or labor
is flexible and satisfies Assumption 3. Notice that it is not necessary that this flexible input be
used as the proxy, just that it satisfies scalar unobservability and monotonicity. Hence, in the
value-added case with intermediate inputs used as the proxy, if labor is also flexible and satisfies
Assumption 3, it will be subject to the same identification problems described in this section. A
notable exception is Ackerberg, Caves, and Frazer (2015), which carefully lays out a description
of the data-generating-processes under which their procedure can be employed for value-added
specifications.12 However, based in part on our non-identification results above, ACF suggest not
applying their procedure to gross output production functions that are not Leontief in intermediate
inputs.
12
In particular, in one of the cases they present, they show that when labor is a flexible input that does not satisfyAssumption 3 due to persistent and unobserved wage shocks, its elasticity is identified.
17
4 The Empirical Content of the Model
Our theorems in the previous section show that the LP/Wooldridge approach is nonparametrically
under-identified. Our next result makes precise the exact source of under-identification. In par-
ticular, if the flexible input elasticity ∂∂mt
f (kt, lt,mt) were nonparametrically known, then the
structure of the second stage of the model is informative enough to identify the remainder of the
production function nonparametrically. We present this result for a single flexible input mt, and
show how it extends to multiple flexible input elasticities in Appendix C.
The idea is that the flexible input elasticity defines a partial differential equation that can be
integrated up to identify the part of the production function f related to the intermediate input
mt.13 By the fundamental theorem of calculus we have
∫∂
∂mt
f (kt, lt,mt) dmt = f (kt, lt,mt) + C (kt, lt) . (7)
Subtracting equation (7) from the production function, and re-arranging terms we have
Yt ≡ yt − εt −∫
∂
∂mt
f (kt, lt,mt) dmt = −C (kt, lt) + ωt. (8)
Notice that Yt is an “observable” random variable as it is a function of data and the flexible input
elasticity which is assumed to be known. It also depends on the ex-post shock εt, which can be
recovered, for example, from the first stages of the OP/LP/ACF procedures.
Applying the Markov structure on productivity that is the basis for the “second stage” moments
of the OP/LP/ACF procedure gives
Yt = −C (kt, lt) + h (Yt−1 + C (kt−1, lt−1)) + ηt. (9)
Since (kt, lt, kt−1, lt−1,mt−1, yt−1 − εt−1) is a transformation of the information set It−1, and
Yt−1 is a function of these variables, we have the orthogonality E [ηt | kt, lt,Yt−1, kt−1, lt−1] = 0
13See Houthakker (1950) for a similar solution to the related problem of how to recover the utility function fromthe demand functions.
18
which implies
E [Yt | kt, lt,Yt−1, kt−1, lt−1] = −C (kt, lt) + h (Yt−1 + C (kt−1, lt−1)) . (10)
This regression identified in the data will allow us to identify the last component of the production
function C up to an additive constant.14,15
We now establish this result formally based on the observations of the above discussion. We
will use the following regularity condition on the support of the regressors
(kt, lt,Yt−1, kt−1, lt−1) (adapted from Newey, Powell, and Vella, 1999).
Assumption 4. For each point(Yt, kt−1, lt−1
)in the support of (Yt−1, kt−1, lt−1), the boundary of
the support of (kt, lt) conditional on(Yt, kt−1, lt−1
)has a probability measure zero.
Assumption 4 is a condition that states that we can independently vary the predetermined inputs
(kt, lt) conditional on (Yt−1, kt−1, lt−1) within the support. This implicitly assumes the existence
of enough variation in the input demand functions for the predetermined inputs to induce open
set variation in them conditional on the lagged output and input values (Yt−1, kt−1, lt−1). This
condition makes explicit the variation that allows for nonparametric identification of the remainder
of the production function under the second stage moments above. A version of this assumption is
thus implicit in the LP/ACF/Wooldridge procedures.
Theorem 3. Under Assumptions 1 - 4, if ∂∂mt
f (kt, lt,mt) is nonparametrically known, then the
production function f is nonparametrically identified up to an additive constant.
Proof. The theorem assumes that ∂∂mt
f (kt, lt,mt) is known almost everywhere in (kt, lt,mt). As-
sumptions 2, 3, and 4 ensure that with probability 1 for any (kt, lt,mt) in the support of the data14Notice that in the OP/LP/ACF procedures, Yt can only be formed for observations in which the proxy vari-
able is strictly positive. Observations that violate the strict monotonicity of the proxy equation need to bedropped from the first stage, which implies that εt cannot be recovered. This introduces a selection bias sinceE [ηt | kt, lt,Yt−1, kt−1, lt−1, ιt > 0] 6= E [ηt | kt, lt,Yt−1, kt−1, lt−1], where ιt is the proxy variable, and equation(10) does not hold. The reason is that firms that receive lower draws of ηt are more likely to choose non-positive valuesof the proxy, and this probability is a function of the other state variables of the firm. An alternative is to form ηt+εt =yt−
∫∂∂mt
f (kt, lt,mt) dmt + C (kt, lt) + h (Yt−1 + C (kt−1, lt−1)) , which does not require one to recover εt, andhence can be formed for all observations. One can then use moments of the formE [ηt + εt | kt, lt,Yt−1, kt−1, lt−1] =0 to recover the rest of the production function: −C (kt, lt) + h (Yt−1 + C (kt−1, lt−1)).
15As it is well known, it is not possible to separately identify a constant in the production function from meanproductivity, E [ωt].
19
there is a set
{(k, l,m) | k = kt, l = lt,m ∈ [m (kt, lt) ,mt]}
also contained in the support for some m (kt, lt). Hence with probability 1 the integral
∫ mt
m(kt,lt)
∂
∂mt
f (kt, lt,m) dm = f (kt, lt,mt) + C (kt, lt)
is identified, where the equality follows from the fundamental theorem of calculus. Therefore, if
two production functions f and f give rise to the same input elasticity ∂∂mt
f (kt, lt,mt) over the
support of the data, then they can only differ by an additive function C (kt, lt) . To identify this
additive function, observe that we can identify the joint distribution of (Yt, kt, lt,Yt−1, kt−1, lt−1)
for Yt defined by (8). Thus the regression function
E [Yt | kt, lt,Yt−1, kt−1, lt−1] = r (kt, lt,Yt−1, kt−1, lt−1) (11)
can be identified for almost all xt = (kt, lt,Yt−1, kt−1, lt−1), where recall that r = −C (kt, lt) +
h (Yt−1 + C (kt−1, lt−1)). Let(C , h
)be a candidate alternative structure. The two structures
(C , h) and(C , h
)are observationally equivalent if and only if
−C (kt, lt) + h (Yt−1 + C (kt−1, lt−1)) = −Ct (kt, lt) + h(Yt−1 + C (kt−1, lt−1)
), (12)
for almost all points in the support of xt. Our support assumption (4) on (kt, lt) allows us to take
partial derivatives of both sides of (12) with respect to kt and lt
∂
∂zC (kt, lt) =
∂
∂zC (kt, lt)
for z ∈ {kt, lt} and for all xt in its support, which implies C (kt, lt)− C (kt, lt) = c for a constant
c for almost all xt. Thus we have shown the production function is identified up to an additive
constant.
The key insight this result provides is that it makes precise the true empirical content of the
20
proxy variable structure places on the economic environment surrounding the production func-
tion. In particular, it shows the empirical content of the structure of the model is such that it can
nonparametrically identify the predetermined input elasticities given the flexible input elasticities.
However, from our results in Section 3, there is not enough variation in the model to identify the
flexible input elasticities themselves. Hence, our Theorems 1 - 3 establish that the precise source
of under-identification of the proxy variable model is the elasticity ∂∂mt
f (kt, lt,mt) of the flexible
input mt that is used as the proxy variable. The conclusion is that the proxy variable structure in
LP/Wooldridge is too weak to identify this flexible input elasticity.
The simple intuition for this fact can be seen by juxtaposing the production function yt =
f (kt, lt,mt) + ωt + εt against the structure on intermediate input demand mt = M (kt, lt, ωt).
Observe first that mt is an endogenous variable in the model as it is also a function of the same
productivity shock ωt that determines output yt. However this endogenous variable does not admit
any source of variation from outside the production function - the only input demand shifter aside
from the other inputs in the production function is productivity ωt. In particular, the elasticity is
identified with how output varies with mt holding fixed (kt, lt), but the only source of variation in
mt (namely ωt) also simultaneously shifts output yt. Thus it would seem impossible to identify an
elasticity of the production function with respect to intermediate inputs mt.
One way out of this problem is to allow for observed shifters that enter the flexible input
demand M, but are excluded from the production function.16 Flexible input and output prices that
vary by firm are one natural source of variation and it has been considered recently by Doraszelski
and Jaumandreu (2013, 2015). In Section 7 we provide a more detailed discussion of the use of
prices as exclusions restrictions, and the circumstances under which they can be used. In the next
section, however, we present an alternative approach to identify the flexible input elasticity which
is widely applicable and does not rely on exogenously varying observable price differences among
firms within an industry.
16It may be possible to achieve identification in the absence of exclusion restrictions by imposing additional re-strictions. One example is using heteroskedasticity restrictions (see e.g., Rigobon, 2003; Klein and Vella, 2010; andLewbel, 2012), although these approaches require explicit restrictions on the form of the error structure. We thank ananonymous referee for pointing this out. We are not aware of any applications of these ideas in the production functionsetting.
21
5 Nonparametric Identification of the Flexible Input Elasticity
The source of the identification problem underlying our Theorems 1 and 2 is that the data cannot
disentangle f from M−1 in either the first or second stages of the proxy variable technique. Our
solution is to use the restrictions implied by optimal firm behavior which underlie the input demand
function M. The key idea is to recognize that f and M are not independent functions for an
optimizing firm, but instead the input demand M is implicitly defined by f through the firm’s first
order condition. We show that this functional relationship can be exploited in a fully nonparametric
fashion (i.e., without imposing parametric structure on f ) to solve the non-identification problem.17
The key reason why we are able to use the first order condition with such generality is that the key
assumption of the model - Assumption 3 - already presumes intermediate inputs are a flexible
input, thus making the economics of this input choice especially tractable.
We focus attention in the main body on the classic case of perfect competition in the interme-
diate input and output markets. The perfect competition case makes our proposed solution to the
identification problem caused by intermediate inputs particularly transparent. In Appendix C4, we
show that our approach can be extended to the case of monopolistic competition with unobserved
output prices.
Assumption 5. Firms are price takers in the output and intermediate input market, with ρt denot-
ing the common intermediate input price and Pt denoting the common output price facing all firms
in period t. The production function f is differentiable at all (k, l,m) ∈ R3++. Firms maximize
expected discounted profits.
We can view Assumption 5 as a natural strengthening of Assumption 3. Indeed the monotonic-
ity of M in Assumption 3 is typically justified following from the market structure, Assumption
5, under suitable shape restrictions on the production function f (see Appendix A in Levinsohn
and Petrin, 2003). However, an important distinction is that our approach can be generalized to
allow for additional unobservables in the firm’s problem that scalar unobservability, Assumption
3, cannot accommodate (see Appendix C).
17Please see Appendix C3 for the extension to the case of multiple flexible inputs.
22
Assumption 5 allows us to establish the link between M and the production function f . The
firm’s profit maximization problem with respect to intermediate inputs is
M (kt, lt, ωt) = arg maxMt
PtE[F (kt, lt,mt) e
ωt+εt | It]− ρtMt, (13)
which follows because Mt does not have any dynamic implications and thus only affects current
period profits. The first order condition of the problem (13) is
Pt∂
∂Mt
F (kt, lt,mt) eωtE = ρt, (14)
where recall E = E [eεt ]. Taking logs of (14) and differencing with the production function (2)
gives
st = ln E + ln
(∂
∂mt
f (kt, lt,mt)
)− εt (15)
≡ lnDE (kt, lt,mt)− εt
where st ≡ ln(ρtMt
PtYt
)is the (log) intermediate input share of output.
Theorem 4. Under Assumptions 1, 2, 4, 5, and that ρt, Pt (or price-deflators) are observed, the
share regression in equation (15) nonparametrically identifies the flexible input elasticity ∂∂mt
f (kt, lt,mt)
of the production function almost everywhere in (kt, lt,mt).
Proof. Because E [εt | kt, lt,mt] = 0 we have that the conditional expectation
E [st | kt, lt,mt] = lnDE (kt, lt,mt) (16)
identifies the function DE . We refer to this regression in the data as the share regression. As we
now show, the share regression is the key to close the non-identification gap left by our Theorems
1 and 2. Observe that εt = lnDE (kt, lt,mt)− st and thus the constant
E = E[exp
(lnDE (kt, lt,mt)− st
)](17)
23
can be identified.18 This allows us to identify the flexible input elasticity as
D (kt, lt,mt) ≡∂
∂mt
f (kt, lt,mt) =DE (kt, lt,mt)
E. (18)
Theorem 4 shows that, by taking full advantage of the economic content of the model, we can
identify the flexible input elasticity. Combined with Theorem 3 this allows us to identify the whole
production function f nonparametrically.
6 A Computationally Simple Estimator
In this section we show how to obtain a simple nonparametric estimator of the production function
using standard sieve series estimators as analyzed by Chen (2007). Our estimation procedure
consists of two steps. We first show how to estimate the share regression, and then proceed to
estimation of the constant of integration C and the Markov process h.
We propose a finite-dimensional truncated linear series given by a complete polynomial of
degree r for the share regression. In what follows, we add back in the firm subscripts j for clarity.
Given the observations {(yjt, kjt, ljt,mjt)}Tt=1 for the firms j = 1, . . . , J sampled in the data, we
propose to use a complete polynomial of degree r in kjt,ljt,mjt and to use the sum of squared
residuals,∑
jt ε2jt, as our objective function. For example, for a complete polynomial of degree
two, our estimator would solve:
minγ′
∑j,t
sjt − ln
γ′0 + γ′kkjt + γ′lljt + γ′mmjt + γ′kkk2jt + γ′lll
2jt
+γ′mmm2jt + γ′klkjtljt + γ′kmkjtmjt + γ′lmljtmjt
2
.
The solution to this problem is an estimator
DEr (kjt, ljt,mjt) =∑
rk+rl+rm≤r
γ′rk,rl,rmkrkjt l
rljtm
rmjt , with rk, rl, rm ≥ 0, (19)
18Doraszelski and Jaumandreu (2013) use a similar idea to recover this constant.
24
of the elasticity up to the constant E , as well as the residual εjt corresponding to the ex-post shocks
to production.19 Since we can estimate E = 1JT
∑j,t e
εjt , we can recover γ ≡ γ′
E , and thus estimate
∂∂mjt
f (kjt, ljt,mjt) from equation (19), free of the constant.
Given our estimator for the intermediate input elasticity, we can calculate the integral in (7).
One advantage of the polynomial sieve estimator we have selected is that this integral will have a
closed-form solution:
Dr (kjt, ljt,mjt) ≡∫Dr (kjt, ljt,mjt) dmjt =
∑rk+rl+rm≤r
γrk,rl,rmrm + 1
krkjt lrljtm
rm+1jt .
For a degree two estimator (r = 2) we would have
D2 (kjt, ljt,mjt) ≡
γ0 + γkkjt + γlljt + γm2mjt + γkkk
2jt + γlll
2jt
+γmm3m2jt + γklkjtljt + γkm
2kjtmjt + γlm
2ljtmjt
mjt.
With an estimate of εjt and of Dr (kjt, ljt,mjt) in hand, we can form a sample analogue of Yjt in
equation (8): Yjt ≡ ln
(Yjt
eεjteDr(kjt,ljt,mjt)
).
In the second step, in order to recover the constant of integration C in (9) and the Markovian
process h, we use similar complete polynomial series estimators. Since a constant in the production
function cannot be separately identified from mean productivity, E [ωjt], we normalize C (kjt, ljt)
to contain no constant. That is, we use
Cτ (kjt, ljt) =∑
0<τk+τl≤τ
ατk,τlkτkjt l
τljt, with τk, τl ≥ 0, (20)
and
hA (ωjt−1) =∑
0≤a≤A
δaωajt−1, with a ≤ A (21)
19As with all nonparametric sieve estimators, the number of terms in the series increases with the number of ob-servations. Under mild regularity conditions these estimators will be consistent and asymptotically normal for sieveM-estimators like the one we propose. See Chen (2007).
25
for some degrees τ and A (that increase with the sample size). Combining these gives us
Yjt = −∑
0<τk+τl≤τ
ατk,τlkτkjt l
τljt +
∑0≤a≤A
δaωajt−1 + ηjt. (22)
Replacing for ωjt−1 we have the estimating equation:
Yjt = −∑
0<τk+τl≤τ
ατk,τlkτkjt l
τljt +
∑0≤a≤A
δa
(Yjt−1 +
∑0<τk+τl≤τ
ατk,τlkτkjt−1l
τljt−1
)a
+ ηjt (23)
We can then use moments of the form E[ηjtk
τkjt l
τljt
]= 0 and E
[ηjtYajt−1
]= 0 to form a
standard sieve GMM criterion function to estimate (α, δ).20 Notice that the estimator described
above is just-identified. One could also use higher-order moments, as well as lags of inputs, to
estimate an over-identified version of the model.
Even though the estimator we introduce is a straightforward application of sieves, to our knowl-
edge, there is no asymptotic distributional theory for multi-step nonparametric sieve estimators.
Hence, for the purpose of inference, we interpret our estimator as a flexible parametric approxima-
tion to the production function. Our estimator then becomes a standard GMM problem that uses
the following moments
E
[εjt∂ lnDr (kjt, ljt,mjt)
∂γ
]= 0,
E[ηjtk
τkjt l
τljt
]= 0,
E[ηjtYajt−1
]= 0,
where the first set of moments are the NLLS moments corresponding to the share equation. We
can then apply standard GMM theory to do inference.21 Since our main empirical focus is not on
20Alternatively, for a guess of α, one can form ωjt−1 (α) = Yjt−1 + C (kjt−1, ljt−1) = Yjt−1 +∑0<τk+τl≤τ ατk,τlk
τkjt−1l
τljt−1, and based on equation (22) use moments of the form E [ηjtωjt−1 (α)] = 0 to esti-
mate δ. Notice that since ωjt (α) = Yjt +∑
0<τk+τl≤τ ατk,τlkτkjt l
τljt, this is equivalent to regressing ωjt on a sieve in
ωjt−1. Then the moments E[ηjtk
τkjt l
τljt
]= 0 can be used to estimate α.
21See e.g., Hansen (1982) for the one-step version and Newey and McFadden (1994) for the two-step version.
26
particular parameters themselves, but on functions of the parameters (e.g., elasticities, productiv-
ity), we employ a nonparametric block bootstrap to compute standard errors (see e.g., Horowitz,
2001).22
7 Relationship to Literature
7.1 Price Variation as an Instrument
Recall that the identification problem with respect to intermediate inputs stems from insufficient
variation in mjt to identify their influence in the production function independently of the other in-
puts. However, by looking at the intermediate input demand equation,mjt = M(kjt, ljt, ωjt, ρt, Pt),
it can be seen that, if prices (Pt, ρt) were firm specific, they could potentially serve as a source of
variation to address the identification problem.23 In fact, given enough sources of price variation,
prices could potentially be used directly as instruments to estimate the entire production function
while controlling for the endogeneity of input decisions.
There are several challenges to using prices as instruments (see GM and Ackerberg et al.,
2007). First, in many firm-level production datasets, firm-specific prices are not observed. Second,
even if price variation is observed, in order to be useful as an instrument, the variation employed
must not be correlated with the innovation to productivity, ηjt (as is likely to be the case if firms
choose prices optimally due to market power); and it cannot purely reflect differences in the quality
of either inputs or output. To the extent that input and output prices capture quality differences,
22Since we estimate a just-identified model, the multi-step and single-step estimators are equivalent. In principle,one could employ the asymptotic results in Chen and Pouzo (2015) and Tao (2015) to do inference for the nonpara-metric estimator in this case. In practice, researchers may want to use an over-identified version of our estimator, inwhich case these results no longer apply. Hence we focus on the flexible parametric interpretation for the purposes ofinference. That being said, in Online Appendix O1, we present Monte Carlo simulations which show that our bootstrapprocedure has the correct coverage for the nonparametric estimates.
23The intermediate input demand equation also suggests that time-varying prices, even if they were not firm-specific,could serve as potential instruments. While industry-time specific price indices (deflators) are commonly available inmost datasets, this approach is problematic for a few reasons. First, these instruments are likely to have little identifyingvariation given that they do not vary across firm. Second, given that most firm-level datasets are short panels, withasymptotics taken in the number of firms as opposed to the number of periods, time-series variation alone will not leadto consistent estimates. In addition, when the production function is allowed to vary over time, time-series variationin prices is no longer sufficient for identification, as variation within each period is necessary.
27
prices should be included in the measure of the inputs used in production.24
This is not to say that if one can isolate exogenous price variation, it cannot be used to aid
in identification. The point is that just observing price variation is not enough. The case must
be made that the price variation that is used is indeed exogenous. For example, if prices are
observed and serially correlated, one way to deal with the endogeneity concern, as suggested by
Doraszelski and Jaumandreu (2013), is to use lagged prices as instruments. This diminishes the
endogeneity concerns, since lagged prices only need to be uncorrelated with the innovation to
productivity, ηjt. In an effort to show that wage variation does not reflect differences in worker
quality, Doraszelski and Jaumandreu (2015) demonstrate empirically that the majority of wage
variation in the Spanish manufacturing dataset they use is not due to variation in the skill mix of
workers, and therefore likely due to geographic and temporal differences in labor markets. This
work demonstrates that prices (specifically lagged prices), when carefully employed, can be a
useful source of variation for identification of the production function. However, as also noted
in Doraszelski and Jaumandreu (2013), this information is not available in most datasets. Our
approach offers an alternative identification strategy that can be employed even when external
instruments are not available.
7.2 Exploiting First-Order Conditions
The use of first-order conditions for the estimation of production functions dates back to at least the
work by Klein (1953) and Solow (1957),25 who recognized that, for a Cobb-Douglas production
function, there is an explicit relationship between the parameters representing input elasticities and
input cost or revenue shares. This observation forms the basis for index number methods (see e.g.,
Caves, Christensen, and Diewert, 1982) that are used to nonparametrically recover input elasticities
and productivity.26
24Recent work has suggested that quality differences may be a key driver of price differences (see GM and Fox andSmeets, 2011).
25Other examples of using first-order conditions to obtain identification include Stone (1954) on consumer demand,Heckman (1974) on labor supply, Hansen and Singleton (1982) on Euler equations and consumption, Paarsch (1992)and Laffont and Vuong (1996) on auctions, and Heckman, Matzkin, and Nesheim (2010) on hedonics.
26Index number methods are grounded in three important economic assumptions. First, all inputs are flexible andcompetitively chosen. Second, the production technology exhibits constant returns to scale, which while not strictly
28
More recently, Doraszelski and Jaumandreu (2013, 2015) and Grieco, Li, and Zhang (2014)
exploit the first-order conditions for labor and intermediate inputs under the assumption that they
are flexibly chosen. Instead of using shares to recover input elasticities, these papers recognize
that given a particular parametric form of the production function, the first-order condition for a
flexible input (the proxy equation in LP/ACF) implies cross-equation parameter restrictions that
can be used to aid in identification. Using a Cobb-Douglas production function, Doraszelski and
Jaumandreu (2013) show that the first-order condition for a flexible input can be re-written to
replace for productivity in the production function. Combined with observed variation in the prices
of labor and intermediate inputs, they are able to estimate the parameters of the production function
and productivity.
Doraszelski and Jaumandreu (2015) extend the methodology developed in Doraszelski and
Jaumandreu (2013) to estimate productivity when it is non-Hicks neutral, for a CES production
function. By exploiting the first order conditions for both labor and intermediate inputs they are
able to estimate a standard Hicks-neutral and a labor-augmenting component to productivity.
Grieco, Li, and Zhang (2014) also use first order conditions for both labor and intermediate
inputs to recover multiple unobservables. In the presence of unobserved heterogeneity in interme-
diate input prices, they show that the parametric cross-equation restrictions between the production
function and the two first-order conditions, combined with observed wages, can be exploited to es-
timate the production function and recover the unobserved intermediate input prices. They also
show that their approach can be extended to account for the composition of intermediate inputs
and the associated (unobserved) component prices.
The paper most closely related to ours is Griliches and Ringstad (1971), which exploits the
relationship between the first order condition for a flexible input and the production function in
a Cobb-Douglas parametric setting. They use the average revenue share of the flexible input to
measure the output elasticity of flexible inputs. This combined with the log-linear form of the
Cobb-Douglas production function, allows them to then subtract out the term involving flexible in-
necessary is typically assumed in order to avoid imputing a rental price of capital. Third, and most importantly forour comparison, there are no ex-post shocks to output. Allowing for ex-post shocks in the index number frameworkcan only be relaxed by assuming that elasticities are constant across firms, i.e., by imposing the parametric structureof Cobb-Douglas.
29
puts. Finally, under the assumption that the non-flexible inputs are predetermined and uncorrelated
with productivity (not just the innovation), they estimate the coefficients for the predetermined
inputs.
Our identification solution can be seen as a nonparametric generalization of the Griliches and
Ringstad (1971) empirical strategy. Instead of using the Cobb-Douglas restriction, our share equa-
tion (15) uses revenue shares to recover input elasticities in a fully nonparametric setting. In
addition, rather than subtract out the effect of intermediate inputs from the production function,
we instead integrate up the intermediate input elasticity and take advantage of the nonparametric
cross-equation restrictions between the share equation and the production function. Furthermore,
we allow for predetermined inputs to be correlated with productivity, but uncorrelated with just the
innovation to productivity.
7.3 Dynamic Panel
An alternative approach employed in the empirical literature is to use the dynamic panel estima-
tors of Arellano and Bond (1991) and Blundell and Bond (1998, 2000). Under a linear parametric
restriction on the evolution of ωt, these methods take advantage of the conditional moment re-
strictions implied by Assumption (2), which allows for the use of appropriately lagged inputs as
instruments.
If one is willing to step outside the model described in Section 2 and assume that all inputs are
not flexible (i.e., rule out the existence of flexible inputs) or that no flexible input satisfies the proxy
variable assumption,27 and in addition assume a version of Assumption 4 that includes all inputs,
then it may be possible to show that these dynamic panel methods can identify the production
function and productivity. However, the bulk of empirical work based on production function
estimation has focused on environments in which some inputs are non-flexible and some inputs
are flexible. It is this setting that motivates our problem and distinguishes our approach from the
27Note that not satisfying the proxy variable assumption does not guarantee identification in the presence of aflexible input. For example, unobserved serially correlated intermediate input price shocks violate the proxy vari-able assumption. However, this variation generates a measurement problem, since intermediate inputs are typicallymeasured in expenditures.
30
dynamic panel literature.
8 Value Added
A common empirical approach is to employ a value-added production function by relating a mea-
sure of the output of a firm to a function of capital and labor only. Typically output is measured
empirically as the “value added” by the firm (i.e., the value of gross output minus expenditures
on intermediate inputs).28 One potential advantage of this approach is that, by excluding inter-
mediate inputs from the production function, it avoids the identification problem associated with
intermediate inputs that we describe in Section 3.
The use of value added is typically motivated in one of two ways. First, a researcher may feel
that a value-added function is a better model of the production process. For example, suppose
there is a lot of heterogeneity in the degree of vertical integration within an industry, with firms
outsourcing varying degrees of the production process, as a result of the production function being
heterogeneous in how intermediate inputs enter. In this case, a researcher may feel that focusing
on just the contributions of capital and labor (to the value added by the firm) is preferred to a
gross output specification including intermediate inputs. Under this setup, if either capital or labor
are flexibly chosen, then our non-identification arguments in Section 3 still apply. Similarly, our
identification arguments also apply, and one can use our proposed identification/estimation strategy
for this model as well.
The second motivation is based on the idea that a value-added function can be constructed from
an underlying gross output production function. This value-added function can then be used to
recover objects of interest from the underlying gross output production function, such as firm-level
productivity eωt+εt and certain features of the production technology (e.g., output elasticities of
inputs) with respect to the “primary inputs”, capital and labor. This approach is typically justified
either via the restricted profit function or by using structural production functions. As we discuss in
more detail below, under the model described in Section 2, in general, neither justification allows
28One exception is Ackerberg, Caves, and Frazer (2015) which uses gross output as the output measure.
31
for a value added production function to be isolated from the gross output production function.
Regardless of the motivation for value added, the objects from a value-added specification,
particularly productivity, will be fundamentally different than those from gross output. Under the
first motivation, this is because productivity from a primitively specified value-added setup mea-
sures differences in value added holding capital and labor fixed, as opposed to differences in gross
output holding all inputs fixed. The results in this section show that under the second motivation,
the value-added objects cannot generally be mapped into their gross output counterparts if all one
has are the value-added objects. A key exception is the linear in intermediate inputs Leontief spec-
ification that we discuss below, a version of which is employed by Ackerberg, Caves, and Frazer
(2015).
8.1 Restricted Profit Value-Added
The first approach to relating gross output to value added is based on the duality results in Bruno
(1978) and Diewert (1978). We first briefly discuss their original results, which were derived under
the assumption that intermediate inputs are flexibly chosen, but excluding the ex-post shocks. In
this case, they show that by replacing intermediate inputs with their optimized value in the profit
function, the empirical measure of value added, V AEt ≡ Yt −Mt can be expressed as:
V AEt = F (kt, lt,M (kt, lt, ωt)) eωt −M (kt, lt, ωt) ≡ V (kt, lt, ωt) , (24)
where we use V (·) to denote value-added in this setup.29 This formulation is sometimes referred
to as the restricted profit function (see Lau, 1976; Bruno, 1978; McFadden, 1978).
In an index number framework, Bruno (1978) shows that elasticities of gross output with re-
spect to capital, labor, and productivity can be locally approximated by multiplying estimates of the
value-added counterparts by the firm-level ratio of value added to gross output, V AEtGOt
= (1− St),
29Technically, V AEt ≡ PtYt
P t− ρtMt
ρt, where P t and ρt are the price deflators for output and intermediate inputs,
respectively. The ratio Pt
P tis equal to the output price in the base year, PBASE , and similarly for the price of interme-
diate inputs. Since PBASE and ρBASE are constants, they get subsumed in the constants in the F and M functions.For ease of notation, we normalize these constants to 1.
32
where GO stands for gross output.30 For productivity, the result is as follows:
(elasGOteωt
)=(elas
V AEteωt
)(V AEtGOt
)=(elas
V AEteωt
)(1− St) (25)
See Online Appendix O2 for the details of this derivation. Analogous results hold for the elasticities
with respect to capital and labor by replacing eωt with Kt or Lt.
While this derivation suggests that estimates from the restricted-profit value-added function can
be simply multiplied by (1− St) to recover estimates from the underlying gross output production
function, there are several important problems with the relationship in equation (25). First, this
approach is based on a local approximation. While this may work well for small changes in
productivity, for example looking at productivity growth rates (the original context under which
these results were derived), it may not work well for large differences in productivity, such as
analyzing cross-sectional productivity differences.
Second, this approximation does not account for ex-post shocks to output. As we show in
Online Appendix O2, when ex-post shocks are accounted for, the relationship in equation (25)
becomes:
(∂GOt
∂eωteωt
GOt
)︸ ︷︷ ︸
elasGOteωt
=
(∂V AEt∂eωt
eωt
V AEt
)︸ ︷︷ ︸
elasV AEteωt
(1− St) +
[∂Mt
∂eωteωt
GOt
(eεt
E− 1
)](26)
The term in brackets is the bias introduced due to the ex-post shock. Ex-post shocks drive a wedge
between the local equivalence of value added and gross output objects. Analogous results for the
output elasticities of capital and labor can be similarly derived.
As a result of the points discussed above, estimates from the restricted profit value-added func-
tion cannot simply be “transformed” by re-scaling with the firm-specific share of intermediate
inputs to obtain estimates of the underlying production function and productivity. How much of a
difference this makes is ultimately an empirical question, which we address in Section 9. Preview-
30These results were originally derived under a general form of technical change. We have augmented the resultshere to correspond to the standard setup with Hicks-neutral technical change as discussed in Section 2.1.
33
ing our results, we find that re-scaling using the shares, as suggested by equation (25), performs
poorly.
8.2 “Structural” Value-Added
The second approach to connecting gross output to value added is based on specific parametric as-
sumptions on the production function, such that a value-added production function of only capital,
labor, and productivity can be both isolated and measured (see Sims, 1969 and Arrow, 1972). We
refer to this version of value added as the “structural value-added production function”.
The empirical literature on value-added production functions often appeals to the extreme case
of perfect complements (i.e., Leontief). A standard representation is:
Yt = min [H (Kt, Lt) , C (Mt)] eωt+εt , (27)
where C (·) is a monotone increasing and concave function. The main idea underlying the Leontief
justification is that, under the assumption that
H (Kt, Lt) = C (Mt) , (28)
the right hand side of equation (27) can be written as H (Kt, Lt) eωt+εt , a function that does not
depend on intermediate inputs M . The key problem with this approach is that, given the assump-
tions of the model, the relation in equation (28) will not generally hold. Unless capital or labor
is assumed to be flexible, firms either cannot adjust them in period t or can only do so with some
positive adjustment cost. The key consequence is that firms may optimally choose to not equate
H(Kt, Lt) and C (Mt), i.e., it may be optimal for the firm to hold onto a larger stock of K and L
than can be combined with M , if K and L are both costly (or impossible) to downwardly adjust.31
31For example, suppose C (Mt) = M0.5t . For simplicity, also suppose that capital and labor are fixed one period
ahead, and therefore cannot be adjusted in the short-run. When M0.5t ≤ H (Kt, Lt), marginal revenue with respect
to intermediate inputs equals ∂C(Mt)∂Mt
aPt. When M0.5t > H (Kt, Lt), increasing Mt does not increase output due to
the Leontief structure, so marginal revenue is zero. Marginal cost in both cases equals the price of intermediate inputs
ρt. The firm’s optimal choice of M is therefore given by Mt =(Pt
ρt0.5a
)2, if(Pt
ρt0.5a
)< H (Kt, Lt). But when
34
An exception to this, as discussed in Ackerberg, Caves, and Frazer (2015), is when C (·) is
linear (i.e., C (Mt) = aMt). In this case the relation in equation (28) will hold, the right hand side
of equation (27) can be written as a function of only capital, labor, and productivity, and we have
that
Yt = H (Kt, Lt) eωt+εt . (29)
This does not imply though that V AE can be used to measure the structural value-added produc-
tion function, as V AEt ≡ Yt −Mt will not be proportional to the value-added production function
H (Kt, Lt) eωt+εt .32 Equation (29), however, suggests that gross output could be used on the left
hand side to measure the structural value-added production function. This is in fact a version of
what Ackerberg, Caves, and Frazer (2015) suggest in estimating a structural value-added produc-
tion function based on an underlying Leontief gross output production function.
9 Data and Application
In the previous section we showed that value-added production functions capture fundamentally
different objects compared to gross output. A natural question that arises is whether these differ-
ences are relevant empirically. A recent survey paper by Syverson (2011) states that many results
in the productivity literature are quite robust to alternative measurement approaches. He attributes
this to the idea that the underlying variation at the firm level is so large that it dominates any dif-
ferences due to measurement. This suggests that whether a researcher uses a value-added or gross
output specification should not change any substantive conclusions related to productivity. In this
section we show that, not only do the two approaches of gross output and value added produce fun-
damentally different patterns of productivity empirically, in many cases the differences are quite
large and lead to very different conclusions regarding the relationship between productivity and
other dimensions of firm heterogeneity.(Pt
ρt0.5a
)> H (Kt, Lt), the firm no longer finds it optimal to set H (Kt, Lt) = C (Mt), and prefers to hold onto
excess capital and labor.32In Online Appendix O2 we also show that moving ωt inside of the min function and/or εt inside of the min
function presents a similar set of issues.
35
We quantify the effect of using a value-added rather than gross output specification using two
commonly employed plant-level manufacturing datasets. The first dataset comes from the Colom-
bian manufacturing census covering all manufacturing plants with more than 10 employees from
1981-1991. This dataset has been used in several studies, including Roberts and Tybout (1997),
Clerides, Lach, and Tybout (1998), and Das, Roberts, and Tybout (2007). The second dataset
comes from the census of Chilean manufacturing plants conducted by Chile’s Instituto Nacional
de Estadística (INE). It covers all firms from 1979-1996 with more than 10 employees. This dataset
has also been used extensively in previous studies, both in the production function estimation liter-
ature (LP) and in the international trade literature (Pavcnik, 2002 and Alvarez and López, 2005).33
We estimate separate production functions for the five largest 3-digit manufacturing industries
in both Colombia and Chile, which are Food Products (311), Textiles (321), Apparel (322), Wood
Products (331), and Fabricated Metal Products (381). We also estimate an aggregate specification
grouping all manufacturing together. We estimate the production function in two ways.34 First, us-
ing our approach from Section 6, we estimate a gross output production function using a complete
polynomial series of degree 2 for both the elasticity and the integration constant in the production
function.35 That is, adding back in the firm subscripts j for clarity, we use
DE2 (kjt, ljt,mjt) = γ′0 + γ′kkjt + γ′lljt + γ′mmjt + γ′kkk2jt + γ′lll
2jt
+γ′mmm2jt + γ′klkjtljt + γ′kmkjtmjt + γ′lmljtmjt
33We construct the variables adopting the convention used by Greenstreet (2007) with the Chilean dataset, andemploy the same approach with the Colombian dataset. In particular, real gross output is measured as deflated rev-enues. Intermediate inputs are formed as the sum of expenditures on raw materials, energy (fuels plus electricity),and services. Real value added is the difference between real gross output and real intermediate inputs, i.e., doubledeflated value added. Labor input is measured as a weighted sum of blue collar and white collar workers, where bluecollar workers are weighted by the ratio of the average blue collar wage to the average white collar wage. Capital isconstructed using the perpetual inventory method where investment in new capital is combined with deflated capitalfrom period t − 1 to form capital in period t. Deflators for Colombia are obtained from Pombo (1999) and deflatorsfor Chile are obtained from Bergoeing, Hernando, and Repetto (2003).
34For all of the estimates we present, we obtain standard errors by using the nonparametric block bootstrap with200 replications.
35We also experimented with higher-order polynomials, and the results were very similar. In a few industries(specifically those with the smallest number of observations) the results are slightly more heterogeneous, as expected.
36
to estimate the intermediate input elasticity and
C2 (kjt, ljt) = αkkjt + αlljt + αkkk2jt + αlll
2jt + αklkjtljt
for the constant of integration. Putting all the elements together, the gross output production func-
tion we estimate is given by:
yjt =
γ0 + γkkjt + γlljt + γm2mjt + γkkk
2jt + γlll
2jt
+γmm3m2jt + γklkjtljt + γkm
2kjtmjt + γlm
2ljtmjt
mjt (30)
−αkkjt − αlljt − αkkk2jt − αlll2jt − αklkjtljt + ωjt + εjt,
since yjt =∫ DE(ljt,kjt,mjt)
E dmjt − C (kjt, ljt) + ωjt + εjt.
Second, we estimate a value-added specification using the commonly-applied method devel-
oped by ACF, also using a complete polynomial series of degree 2:
vajt = βkkjt + βlljt + βkkk2jt + βlll
2jt + βklkjtljt + υjt + εjt, (31)
where υjt + εjt represents productivity in the value-added model.
In Table 1 we report estimates of the average output elasticities for each input, as well as
the sum, for both the value-added and gross output models.36 In every case but one, the value-
added model generates a sum of elasticities that is larger relative to gross output, with an average
difference of 2% in Colombia and 6% in Chile.
We also report the ratio of the mean capital and labor elasticities, which measures the capital in-
tensity (relative to labor) of the production technology in each industry. In general, the value-added
estimates of the capital intensity of the technology are larger relative to gross output, although the
differences are small. According to both measures, the Food Products (311) and Textiles (321)
industries are the most capital intensive in Colombia, and in Chile the most capital intensive are
36The distributions of the estimated firm-specific output elasticities are quite reasonable. For each industry, less than2% are outside of the range (0,1) for labor and intermediate inputs. For capital, the elasticities are closer to zero, buteven in the worst case, less than 9.4% have values below zero. Not surprisingly, these percentages are highest amongthe the industries with the smallest number of observations.
37
Food Products (311), Textiles (321), and Fabricated Metals (381). In both countries, Apparel (322)
and Wood Products (331) are the least capital intensive industries, even compared to the aggregate
specification denoted “All” in the tables.
Value added also recovers dramatically different patterns of productivity as compared to gross
output. Following OP, we define productivity (in levels) as the sum of the persistent and unantici-
pated components: eω+ε.37 In Table 2 we report estimates of several frequently analyzed statistics
of the resulting productivity distributions. In the first three rows of each panel we report ratios
of percentiles of the productivity distribution, a commonly used measure of productivity disper-
sion. There are two important implications of these results. First, value added suggests a much
larger amount of heterogeneity in productivity across plants within an industry, as the various per-
centile ratios are much smaller under gross output. For Colombia, the average 75/25, 90/10, and
95/5 ratios are 1.88, 3.69, and 6.41 under value added, and 1.33, 1.78, and 2.23 under gross out-
put. For Chile, the average 75/25, 90/10, and 95/5 ratios are 2.76, 8.02, and 17.93 under value
added, and 1.48, 2.20, and 2.95 under gross output. The value-added estimates imply that, with the
same amount of inputs, the 95th percentile plant would produce more than 6 times more output in
Colombia, and almost 18 times more output in Chile, than the 5th percentile plant. In stark con-
trast, we find that under gross output, the 95th percentile plant would produce only 2 times more
output in Colombia, and 3 times more output in Chile, than the 5th percentile plant with the same
inputs.
In addition, the ranking of industries according to the degree of productivity dispersion is not
preserved moving from the value added to gross output estimates. For example, in Chile, the
Fabricated Metals industry (381) is found to have the smallest amount of productivity dispersion
under value added, but the largest amount of dispersion under gross output, for all three dispersion
measures.
The second important result is that value added also implies much more heterogeneity across
37Since our interest is in analyzing productivity heterogeneity we conduct our analysis using productivity in levels.An alternative would be to measure productivity in logs. However, the log transformation is only a good approximationfor measuring percentage differences in productivity across groups when these differences are small, which they arenot in our data. We have also computed results based on log productivity. As expected, the magnitude of our resultschanges, however, our qualitative results comparing gross output and value added still hold. We have also computedresults using just the persistent component of productivity, eω . The results are qualitatively similar.
38
industries, which is captured by the finding that the range of the percentile ratios across industries
is much tighter using the gross output measure of productivity. For example, for the 95/5 ratio, the
value-added estimates indicate a range from 4.36 to 11.01 in Colombia and from 12.52 to 25.08
in Chile, whereas the gross output estimates indicate a range from 2.02 to 2.38 and from 2.48 to
3.31. The surprising aspect of these results is that the dispersion in productivity appears far more
stable both across industries and across countries when measured via gross output as opposed to
value added. In the conclusion we sketch some important policy implications of this finding for
empirical work on the misallocation of resources.
In addition to showing much larger overall productivity dispersion, results based on value
added also suggest a substantially different relationship between productivity and other dimen-
sions of plant-level heterogeneity. We examine several commonly-studied relationships between
productivity and other plant characteristics. In the last four rows of each panel in Table 2 we report
percentage differences in productivity based on whether plants export some of their output, import
intermediate inputs, have positive advertising expenditures, and pay above the median (industry)
level of wages.
Using the value-added estimates, for most industries exporters are found to be more productive
than non-exporters, with exporters appearing to be 83% more productive in Colombia and 14%
more productive in Chile across all industries. Using the gross output specification, these estimates
of productivity differences fall to 9% in Colombia and 3% in Chile, and actually turn negative
(although not statistically different from zero) in some cases.
A similar pattern exists when looking at importers of intermediate inputs. The average pro-
ductivity difference is 14% in Colombia and 41% in Chile using value added. However, under
gross output, these numbers fall to 8% and 13% respectively. The same story holds for differences
in productivity based on advertising expenditures. Moving from value added to gross output, the
estimated difference in productivity drops for most industries in Colombia, and for all industries
in Chile. In several cases it becomes statistically indistinguishable from zero.
Another striking contrast arises when we compare productivity between plants that pay wages
above versus below the industry median. Using the productivity estimates from a value-added
39
specification, firms that pay wages above the median industry wage are found to be substantially
more productive, with the estimated differences ranging from 34%-63% in Colombia and from
47%-123% in Chile. In every case the estimates are statistically significant. Using the gross output
specification, these estimates fall to 9%-22% in Colombia and 19%-30% in Chile, representing a
fall by a factor of 3, on average, in both countries.
Since intermediate input usage is likely to be positively correlated with productivity, we would
expect that including (excluding) intermediate inputs in the production function will lead to smaller
(larger) differences in productivity heterogeneity. Therefore, we would expect to see the largest
discrepancies between the value-added and gross output productivity heterogeneity estimates in
industries which are intensive in intermediate input usage. By looking at Tables 1 and 2 we can
confirm that, for the most part, this is the case. When comparing the value added and gross output
productivity estimates, the largest differences tend to occur in the most intermediate input inten-
sive industries, which are Food Products (311) in Colombia and Food Products (311) and Wood
Products (331) in Chile. However, this is not always the case. For example, in Chile, the differ-
ence between the gross output and value added estimates of the average productivity comparing
advertisers and non-advertisers is actually the smallest in the Wood Products (331) industry.
In order to isolate the importance of the value-added/gross output distinction separately from
the effect of transmission bias, in Table 3 we repeat the above analysis without correcting for the
endogeneity of inputs. We examine the raw effects in the data by estimating productivity using
simple linear regression (OLS) to estimate both gross output and value-added specifications, using
a complete polynomial of degree 2. As can be seen from Table 3, the general pattern of results,
that value added leads to larger productivity differences across many dimensions, is similar to our
previous results both qualitatively and quantitatively.
While the results in Table 3 may suggest that transmission bias is not empirically important,
in Table 4 we show evidence to the contrary. In particular, we report the average input elasticities
based on estimates for the gross output model using OLS and using our method to correct for
transmission bias. A well-known result is that failing to control for transmission bias leads to
overestimates of the coefficients on more flexible inputs. The intuition is that the more flexible
40
the input is, the more it responds to productivity shocks and the higher the degree of correlation
between that input and unobserved productivity. The estimates in Table 4 show that the OLS results
substantially overestimate the output elasticity of intermediate inputs in every case. The average
difference is 34%, which illustrates the importance of controlling for the endogeneity generated by
the correlation between input decisions and productivity.
An important implication of our results is that, while controlling for transmission bias certainly
has an effect, the use of value added versus gross output has a much larger effect on the productivity
estimates than transmission bias. This suggests that the use of gross output versus value added may
be more important from a policy perspective than controlling for the transmission bias that has
been the primary focus in the production function literature. Our approach enables the estimation
of gross output production functions by solving the identification problem associated with flexible
inputs while simultaneously correcting for transmission bias.
9.1 Robustness Checks
Adjusting the Value Added Estimates As discussed in Section 8.1, in the absence of ex-post
shocks, the derivation provided in equation (25) suggests that the differences between gross output
and value added can be eliminated by re-scaling the value-added estimates by a factor equal to the
plant-level ratio of value added to gross output, i.e., one minus the share of intermediate inputs
in total output. While this idea has been known in the literature for a while, this re-scaling is
very rarely applied in practice.38 As shown in Section 8.1, there are several reasons why this re-
scaling may not work. In order to investigate how well the re-scaling of value added estimates
performs, we apply the transformation implied by equation (25) using the firm-specific ratio of
value added to gross output(V AEtGOt
), a quantity readily available in the data. We find that this re-
scaling performs quite poorly in recovering the underlying gross output estimates of the production
function and productivity, leading to estimates that are in some cases even further from the gross
output estimates than the value-added estimates themselves.
In Tables 5 and 6 we report the re-scaled estimates as well as the value-added estimates using
38See Petrin and Sivadasan (2013) for an example in which a version of this is implemented.
41
ACF and the gross output estimates using our method for comparison. At first glance, the re-
scaling appears to be working as many of the re-scaled value-added estimates move towards the
gross output estimates. However, in some cases, the estimates of dispersion and the relationship
between productivity and other dimensions of firm heterogeneity move only slightly towards the
gross output estimates, and remain very close to the original value-added estimates. Moreover,
in many cases the estimates overshoot the gross output estimates. Even worse, in some cases the
re-scaling moves them in the opposite direction and leads to estimates that are even further from
the gross output estimates than the original value-added estimates. Finally, in several cases, the
re-scaled estimates actually lead to a sign-reversal compared to both the value-added and gross
output estimates. Overall, while in some cases the re-scaling applied to the value-added estimates
moves them closer to the gross output estimates, it does a poor job of replicating the gross output
estimates, and in many cases moves them even further away.
Alternative Assumptions on Flexible Inputs Our new identification and estimation strategy
takes advantage of the first-order condition with respect to a flexible input. We have used interme-
diate inputs (the sum of raw materials, energy, and services) as the flexible input, as intermediate
inputs have been commonly assumed to be flexible in the literature. We believe that this is a rea-
sonable assumption because a) the model period is typically a year and b) what is required is that
they can be adjusted flexibly at the margin. To the extent that spot markets for commodities exist,
including energy and certain raw materials, this enables firms to make such adjustments. However,
it may be the case that in some applications, researchers do not want to assume that all intermediate
inputs are flexible, or they may want to test the sensitivity of their estimates to this assumption.
First, in order to investigate the robustness of our estimator, in Online Appendix O1 we present
results from a Monte Carlo experiment in which we introduce adjustment frictions for the flexible
input. We then estimate the production function using our estimator, assuming the first order
condition holds. We design the simulation so that dynamic panel methods should work well in the
presence of adjustment costs, and compare the performance of our method to a version of dynamic
panel.
42
We find that our method performs remarkably well overall, even for large values of adjustment
costs. As expected, when there are no adjustment costs, our method recovers the true elasticities
of the production function, and dynamic panel breaks down. As we increase adjustment costs, our
method continues to outperform dynamic panel for small values, performs similarly for intermedi-
ate values, and does only marginally worse for large values.
Second, as a robustness check on our results, we estimate two different specifications of our
model in which we allow some of the components of intermediate inputs to be non-flexible. In
particular, the production function we estimate is of the form F (kt, lt, rmt, nst) eωt+εt , where rm
denotes raw materials and ns denotes energy plus services. In one specification we assume rm
to be non-flexible and ns to be flexible, and in the other specification we assume the opposite.
See Appendix D for these results. Overall the results are sensible and qualitatively similar to our
main results. In addition, the relationship between productivity estimates based on gross output
and value added is very similar to the one in the main set of results.
Fixed Effects As we detail in Appendix C, our identification and estimation strategy can be
easily extended to incorporate fixed effects in the production function. The production function
allowing for fixed effects, a, can be written as Yt = F (kt, lt,mt) ea+ωt+εt .39 A common drawback
of models with fixed effects is that the differencing of the data needed to subtract out the fixed
effects can remove a large portion of the identifying information in the data. In the context of
production functions, this often leads to estimates of the capital coefficient and returns to scale that
are unrealistically low, as well as large standard errors (see GM).
In Appendix D, we report estimates corresponding to those in Tables 1 and 2, using our method
to estimate the gross output production function allowing for fixed effects. The elasticity estimates
for intermediate inputs are exactly the same as in the specification without fixed effects, as the first
stage of our approach does not depend on the presence of fixed effects. We do find some evidence
in Colombia of the problems mentioned above as the sample sizes are smaller than those for Chile.
Despite this, the estimates are very similar to those from the main specification for both countries,
39See Kasahara, Schrimpf, and Suzuki (2015) for an important extension of our approach to the general case offirm-specific production functions.
43
and the larger differences are associated with larger standard errors.
Extra Unobservables As we show in Appendix C, our approach can also be extended to incor-
porate additional unobservables driving the intermediate input demand. Specifically, we allow for
an additional unobservable in the share equation for the flexible input (e.g., optimization error).
This introduces some small changes to the identification and estimation procedure, but the core
ideas are unchanged. In Appendix D we report estimates from this alternative specification. Our
results are remarkably robust. The standard errors increase slightly, which is not surprising given
that we have introduced an additional unobservable into the model. The point estimates, however,
are very similar.
10 Conclusion
In this paper we show that the nonparametric identification of production functions in the presence
of both flexible and non-flexible inputs has remained an unresolved issue. We offer a new iden-
tification strategy that closes this loop. The key to our approach is exploiting the nonparametric
cross-equation restrictions between first order condition for the flexible inputs and the production
function.
Our empirical analysis demonstrates that value added can generate substantially different pat-
terns of productivity heterogeneity as compared to gross output, which suggests that empirical
studies of productivity based on value added may lead to fundamentally different policy impli-
cations compared to those based on gross output. To illustrate this possibility, consider the recent
literature that uses productivity dispersion to explain cross-country differences in output per worker
through resource misallocation. As an example, the recent influential paper by Hsieh and Klenow
(2009) finds substantial heterogeneity in productivity dispersion (defined as the variance of log
productivity) across countries as measured using value added. In particular, when they compare
the United States with China and India, the variance of log productivity ranges from 0.40-0.55
for China and 0.45-0.48 for India, but only from 0.17-0.24 for the United States. They then use
44
this estimated dispersion to measure the degree of misallocation of resources in the respective
economies. In their main counterfactual they find that, by reducing the degree of misallocation in
China and India to that of the United States, aggregate TFP would increase by 30%-50% in China
and 40%-60% in India. In our datasets for Colombia and Chile the corresponding estimates of
the variance in log productivity using a value-added specification are 0.43 and 0.94, respectively.
Thus their analysis applied to our data would suggest that there is similar room for improvement
in aggregate TFP in Colombia, and much more in Chile.
However, when productivity is measured using our gross output framework, our empirical
findings suggest a much different result. The variance of log productivity using gross output is
0.08 in Colombia and 0.15 in Chile. These significantly smaller dispersion measures could imply
that there is much less room for improvement in aggregate productivity for Colombia and Chile.
Since the 90/10 ratios we obtain for Colombia and Chile using gross output are quantitatively very
similar to the estimates obtained by Syverson (2004) for the United States (who also employed
gross output but in an index number framework), this also suggests that the degree of differences
in misallocation of resources between developed and developing countries may not be as large as
the analysis of Hsieh and Klenow (2009) implies.40
Exploring the role of gross output production functions for policy problems such as the one
above could be a fruitful direction for future research. A key message of this paper is that insights
derived under value added, compared to gross output, could lead to significantly different policy
conclusions. Our identification strategy provides researchers with a stronger foundation for using
gross output production functions in practice.
40Hsieh and Klenow note that their estimate of log productivity dispersion for the United States is larger thanprevious estimates by Foster, Haltiwanger, and Syverson (2008) by a factor of almost 4. They attribute this to thefact that Foster, Haltiwanger, and Syverson use a selected set of homogeneous industries. However, another importantdifference is that Foster, Haltiwanger, and Syverson use gross output measures of productivity rather than value-addedmeasures. Given our results in Section 9, it is likely that a large part of this difference is due to Hsieh and Klenow’suse of value added, rather than their selection of industries.
45
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50
Colombia
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Labor 0.70 0.22 0.65 0.32 0.83 0.42 0.86 0.44 0.89 0.43 0.78 0.35
(0.04) (0.02) (0.06) (0.03) (0.03) (0.02) (0.06) (0.05) (0.04) (0.02) (0.01) (0.01)
Capital 0.33 0.12 0.36 0.16 0.16 0.05 0.12 0.04 0.25 0.10 0.31 0.14
(0.02) (0.01) (0.04) (0.02) (0.02) (0.01) (0.04) (0.02) (0.03) (0.01) (0.01) (0.01)
Intermediates -- 0.67 -- 0.54 -- 0.52 -- 0.51 -- 0.53 -- 0.54
(0.01) (0.01) (0.01) (0.01) (0.01) (0.00)
Sum 1.03 1.01 1.01 1.01 0.99 0.99 0.98 0.99 1.14 1.06 1.09 1.04
(0.03) (0.01) (0.04) (0.02) (0.02) (0.01) (0.07) (0.04) (0.02) (0.01) (0.01) (0.00)
Mean(Capital) /
Mean(Labor) 0.47 0.55 0.55 0.49 0.19 0.12 0.14 0.08 0.28 0.23 0.39 0.40
(0.06) (0.08) (0.10) (0.09) (0.03) (0.04) (0.05) (0.05) (0.04) (0.04) (0.02) (0.03)
Chile
Labor 0.77 0.28 0.93 0.45 0.95 0.45 0.92 0.40 0.96 0.52 0.77 0.38
(0.02) (0.01) (0.04) (0.03) (0.04) (0.02) (0.04) (0.02) (0.04) (0.03) (0.01) (0.01)
Capital 0.33 0.11 0.24 0.11 0.20 0.06 0.19 0.07 0.25 0.13 0.37 0.16
(0.01) (0.01) (0.02) (0.01) (0.03) (0.01) (0.02) (0.01) (0.02) (0.01) (0.01) (0.00)
Intermediates -- 0.67 -- 0.54 -- 0.56 -- 0.59 -- 0.50 -- 0.55
(0.00) (0.01) (0.01) (0.01) (0.01) (0.00)
Sum 1.10 1.05 1.17 1.10 1.14 1.08 1.11 1.06 1.22 1.15 1.13 1.09
(0.02) (0.01) (0.03) (0.02) (0.03) (0.02) (0.03) (0.01) (0.03) (0.02) (0.01) (0.01)
Mean(Capital) /
Mean(Labor) 0.43 0.39 0.26 0.24 0.21 0.14 0.21 0.18 0.26 0.25 0.48 0.43
(0.03) (0.03) (0.03) (0.04) (0.04) (0.03) (0.03) (0.03) (0.03) (0.03) (0.02) (0.02)
Notes:
Fabricated Metals
(381) All
(Structural Estimates: Value Added vs. Gross Ouput)
d. The row titled "Sum" reports the sum of the average labor, capital, and intermediate input elasticities, and the row titled "Mean(Capital)/Mean(Labor)" reports the ratio of the average capital elasticity to the average labor elasticity.
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers in the first column are based on a value-added specification and are estimated using a complete polynomial series of degree 2 with the method from Ackerberg, Caves, and Frazer (2006). The numbers in the second column are based on a gross
output specification and are estimated using a complete polynomial series of degree 2 for each of the two nonparametric functions (G and C ) of our approach.
c. Since the input elasticities are heterogeneous across firms, we report the average input elasticities within each given industry.
Table 1: Average Input Elasticities of Output
Industry (ISIC Code)
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
51
Colombia
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
Value
Added
(ACF)
Gross
Output
(GNR)
75/25 ratio 2.20 1.33 1.97 1.35 1.66 1.29 1.73 1.30 1.78 1.31 1.95 1.37(0.07) (0.02) (0.09) (0.03) (0.03) (0.01) (0.08) (0.04) (0.04) (0.02) (0.17) (0.01)
90/10 ratio 5.17 1.77 3.71 1.83 2.87 1.66 3.08 1.80 3.33 1.74 4.01 1.86(0.27) (0.05) (0.30) (0.07) (0.09) (0.03) (0.38) (0.12) (0.13) (0.03) (0.07) (0.02)
95/5 ratio 11.01 2.24 6.36 2.38 4.36 2.02 4.58 2.24 5.31 2.16 6.86 2.36(1.11) (0.08) (0.76) (0.14) (0.22) (0.05) (1.01) (0.22) (0.34) (0.06) (0.02) (0.03)
Exporter 3.62 0.14 0.20 0.02 0.16 0.05 0.26 0.15 0.20 0.08 0.51 0.06(0.99) (0.05) (0.10) (0.03) (0.07) (0.03) (0.63) (0.14) (0.05) (0.03) (0.12) (0.01)
Importer -0.25 0.04 0.27 0.05 0.29 0.12 0.06 0.05 0.26 0.10 0.20 0.11(0.08) (0.02) (0.10) (0.04) (0.08) (0.03) (0.53) (0.08) (0.06) (0.02) (0.05) (0.01)
Advertiser -0.46 -0.03 0.20 0.08 0.13 0.05 0.02 0.04 0.15 0.05 -0.13 0.03(0.10) (0.02) (0.07) (0.03) (0.04) (0.02) (0.09) (0.04) (0.04) (0.02) (0.06) (0.01)
Wages > Median 0.59 0.09 0.60 0.18 0.41 0.18 0.34 0.15 0.55 0.22 0.63 0.20(0.19) (0.02) (0.09) (0.03) (0.03) (0.02) (0.17) (0.04) (0.06) (0.02) (0.05) (0.01)
Chile
75/25 ratio 2.92 1.37 2.56 1.48 2.58 1.43 3.06 1.50 2.45 1.53 3.00 1.55(0.05) (0.01) (0.07) (0.02) (0.07) (0.02) (0.08) (0.02) (0.06) (0.02) (0.03) (0.01)
90/10 ratio 9.02 1.90 6.77 2.16 6.76 2.11 10.12 2.32 6.27 2.33 9.19 2.39(0.30) (0.02) (0.30) (0.05) (0.33) (0.05) (0.60) (0.05) (0.27) (0.05) (0.15) (0.02)
95/5 ratio 21.29 2.48 13.56 2.91 14.21 2.77 25.08 3.11 12.52 3.13 20.90 3.31(0.99) (0.05) (0.84) (0.09) (0.77) (0.09) (2.05) (0.11) (0.78) (0.10) (0.47) (0.04)
Exporter 0.27 0.02 0.07 0.02 0.18 0.09 0.12 0.00 0.03 -0.01 0.20 0.03(0.10) (0.02) (0.07) (0.03) (0.08) (0.03) (0.12) (0.03) (0.06) (0.03) (0.04) (0.01)
Importer 0.71 0.14 0.22 0.10 0.31 0.14 0.44 0.15 0.30 0.11 0.46 0.15(0.11) (0.02) (0.05) (0.02) (0.05) (0.02) (0.10) (0.03) (0.05) (0.02) (0.03) (0.01)
Advertiser 0.18 0.04 0.09 0.04 0.15 0.06 0.04 0.03 0.07 0.01 0.14 0.06(0.05) (0.01) (0.04) (0.02) (0.04) (0.02) (0.04) (0.01) (0.04) (0.02) (0.02) (0.01)
Wages > Median 1.23 0.21 0.47 0.19 0.62 0.22 0.68 0.21 0.56 0.22 0.99 0.30(0.09) (0.01) (0.06) (0.02) (0.06) (0.02) (0.08) (0.02) (0.06) (0.02) (0.04) (0.01)
Notes:
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers in the first column are based on a value-added specification and are estimated using a complete polynomial series of degree 2 with the method from Ackerberg, Caves, and Frazer (2006). The numbers in the second column are based on
a gross output specification and are estimated using a complete polynomial series of degree 2 for each of the nonparametric functions (G and C ) of our approach.
c. In the first three rows we report ratios of productivity for plants at various percentiles of the productivity distribution. In the remaining four rows we report estimates of the productivity differences between plants (as a fraction) based on whether they have exported
some of their output, imported intermediate inputs, spent money on advertising, and paid wages above the industry median. For example, in industry 311 for Chile value added implies that a firm that advertises is, on average, 18% more productive than a firm that does
not advertise.
Wood Products
(331)
Fabricated Metals
(381) All
(Structural Estimates)
Table 2: Heterogeneity in Productivity
Industry (ISIC Code)
Food Products
(311)
Textiles
(321)
Apparel
(322)
52
Colombia
Value
Added
(OLS)
Gross
Output
(OLS)
Value
Added
(OLS)
Gross
Output
(OLS)
Value
Added
(OLS)
Gross
Output
(OLS)
Value
Added
(OLS)
Gross
Output
(OLS)
Value
Added
(OLS)
Gross
Output
(OLS)
Value
Added
(OLS)
Gross
Output
(OLS)
75/25 ratio 2.17 1.16 1.86 1.21 1.65 1.17 1.72 1.23 1.78 1.23 1.93 1.24(0.06) (0.01) (0.06) (0.01) (0.03) (0.01) (0.06) (0.02) (0.04) (0.01) (0.02) (0.00)
90/10 ratio 5.15 1.42 3.50 1.51 2.81 1.44 3.05 1.57 3.30 1.53 3.96 1.58(0.27) (0.02) (0.18) (0.04) (0.08) (0.02) (0.22) (0.06) (0.12) (0.02) (0.06) (0.01)
95/5 ratio 10.86 1.74 5.77 1.82 4.23 1.74 4.67 2.01 5.22 1.82 6.81 1.94(0.94) (0.05) (0.55) (0.08) (0.20) (0.04) (0.72) (0.15) (0.31) (0.04) (0.15) (0.02)
Exporter 3.42 0.09 -0.03 -0.01 0.10 0.00 0.21 0.10 0.12 0.03 0.45 0.01(0.99) (0.04) (0.04) (0.01) (0.05) (0.01) (0.19) (0.09) (0.04) (0.02) (0.12) (0.01)
Importer -0.23 -0.02 0.09 0.00 0.21 0.02 0.02 -0.03 0.20 0.05 0.14 0.04(0.07) (0.01) (0.06) (0.01) (0.06) (0.01) (0.06) (0.02) (0.05) (0.01) (0.04) (0.01)
Advertiser -0.46 -0.07 0.11 -0.04 0.10 -0.03 0.01 -0.02 0.08 0.00 -0.16 -0.02(0.11) (0.02) (0.05) (0.02) (0.03) (0.01) (0.07) (0.03) (0.04) (0.01) (0.06) (0.01)
Wages > Median 0.51 0.06 0.49 0.10 0.39 0.13 0.33 0.11 0.50 0.13 0.56 0.13(0.15) (0.02) (0.07) (0.02) (0.03) (0.01) (0.08) (0.03) (0.04) (0.01) (0.05) (0.01)
Chile
75/25 ratio 2.91 1.30 2.57 1.40 2.56 1.36 3.07 1.39 2.47 1.46 3.01 1.45(0.05) (0.00) (0.07) (0.01) (0.07) (0.01) (0.08) (0.01) (0.06) (0.01) (0.03) (0.00)
90/10 ratio 9.00 1.72 6.63 1.97 6.64 1.91 10.21 2.03 6.27 2.14 9.13 2.14(0.29) (0.01) (0.31) (0.04) (0.29) (0.03) (0.57) (0.04) (0.26) (0.04) (0.15) (0.01)
95/5 ratio 20.93 2.15 13.49 2.57 14.20 2.45 25.26 2.77 12.18 2.80 20.64 2.86(0.96) (0.02) (0.83) (0.07) (0.80) (0.05) (2.05) (0.07) (0.77) (0.06) (0.47) (0.03)
Exporter 0.17 -0.01 0.04 -0.02 0.12 0.01 0.12 -0.02 0.00 0.00 0.15 -0.01(0.09) (0.02) (0.06) (0.02) (0.08) (0.02) (0.09) (0.02) (0.06) (0.02) (0.04) (0.01)
Importer 0.57 0.03 0.20 0.04 0.26 0.06 0.41 0.07 0.27 0.06 0.41 0.09(0.09) (0.01) (0.04) (0.02) (0.05) (0.01) (0.09) (0.03) (0.05) (0.02) (0.03) (0.01)
Advertiser 0.12 0.00 0.07 0.01 0.11 0.02 0.02 0.01 0.05 0.01 0.10 0.04(0.04) (0.01) (0.04) (0.01) (0.04) (0.01) (0.04) (0.01) (0.04) (0.02) (0.02) (0.01)
Wages > Median 1.11 0.12 0.45 0.15 0.58 0.16 0.66 0.13 0.53 0.16 0.94 0.24(0.07) (0.01) (0.05) (0.02) (0.06) (0.02) (0.07) (0.02) (0.06) (0.02) (0.03) (0.01)
Notes:
Industry (ISIC Code)
(Uncorrected OLS Estimates)
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers in the first column are based on a value-added specification and are estimated using a complete polynomial series of degree 2 with OLS. The numbers in the second column are based on a gross output specification estimated using a
complete polynomial series of degree 2 with OLS.
c. In the first three rows we report ratios of productivity for plants at various percentiles of the productivity distribution. In the remaining four rows we report estimates of the productivity differences between plants (as a fraction) based on whether they have exported
some of their output, imported intermediate inputs, spent money on advertising, and paid wages above the industry median. For example, in industry 311 for Chile value added implies that a firm that advertises is, on average, 12% more productive than a firm that does
not advertise.
All
Table 3: Heterogeneity in Productivity
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381)
53
Colombia
Gross
Output
(OLS)
Gross
Output
(GNR)
Gross
Output
(OLS)
Gross
Output
(GNR)
Gross
Output
(OLS)
Gross
Output
(GNR)
Gross
Output
(OLS)
Gross
Output
(GNR)
Gross
Output
(OLS)
Gross
Output
(GNR)
Gross
Output
(OLS)
Gross
Output
(GNR)
Labor 0.15 0.22 0.21 0.32 0.32 0.42 0.32 0.44 0.29 0.43 0.26 0.35
(0.01) (0.02) (0.02) (0.03) (0.01) (0.02) (0.03) (0.05) (0.02) (0.02) (0.01) (0.01)
Capital 0.04 0.12 0.06 0.16 0.01 0.05 0.03 0.04 0.03 0.10 0.06 0.14
(0.01) (0.01) (0.01) (0.02) (0.01) (0.01) (0.01) (0.02) (0.01) (0.01) (0.00) (0.01)
Intermediates 0.82 0.67 0.76 0.54 0.68 0.52 0.65 0.51 0.73 0.53 0.72 0.54
(0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.02) (0.01) (0.01) (0.01) (0.00) (0.00)
Sum 1.01 1.01 1.03 1.01 1.01 0.99 1.00 0.99 1.05 1.06 1.04 1.04
(0.01) (0.01) (0.01) (0.02) (0.01) (0.01) (0.02) (0.04) (0.01) (0.01) (0.00) (0.00)
Mean(Capital) /
Mean(Labor) 0.27 0.55 0.27 0.49 0.04 0.12 0.08 0.08 0.11 0.23 0.23 0.40
(0.07) (0.08) (0.06) (0.09) (0.02) (0.04) (0.05) (0.05) (0.04) (0.04) (0.01) (0.03)
Chile
Labor 0.17 0.28 0.26 0.45 0.29 0.45 0.20 0.40 0.32 0.52 0.20 0.38
(0.01) (0.01) (0.02) (0.03) (0.02) (0.02) (0.01) (0.02) (0.02) (0.03) (0.01) (0.01)
Capital 0.05 0.11 0.06 0.11 0.03 0.06 0.02 0.07 0.07 0.13 0.09 0.16
(0.00) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.00) (0.00)
Intermediates 0.83 0.67 0.75 0.54 0.74 0.56 0.81 0.59 0.71 0.50 0.77 0.55
(0.01) (0.00) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.01) (0.00) (0.00)
Sum 1.05 1.05 1.06 1.10 1.06 1.08 1.04 1.06 1.10 1.15 1.06 1.09
(0.00) (0.01) (0.01) (0.02) (0.01) (0.02) (0.01) (0.01) (0.01) (0.02) (0.00) (0.01)
Mean(Capital) /
Mean(Labor) 0.28 0.39 0.22 0.24 0.12 0.14 0.12 0.18 0.21 0.25 0.42 0.43
(0.03) (0.03) (0.04) (0.04) (0.03) (0.03) (0.05) (0.03) (0.04) (0.03) (0.02) (0.02)
Notes:
Fabricated Metals
(381) All
(Gross Output: Structural vs. Uncorrected OLS Estimates)
d. The row titled "Sum" reports the sum of the average labor, capital, and intermediate input elasticities, and the row titled "Mean(Capital)/Mean(Labor)" reports the ratio of the average capital elasticity to the average labor elasticity.
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
c. Since the input elasticities are heterogeneous across firms, we report the average input elasticities within each given industry.
b. For each industry, the numbers in the first column are based on a gross output specification and are estimated using a complete polynomial series of degree 2 with OLS. The numbers in the second column are also based on a gross output specification using a complete
polynomial series of degree 2 for each of the two nonparametric functions (G and C ) of our approach.
Table 4: Average Input Elasticities of Output
Industry (ISIC Code)
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
54
Colo
mb
ia
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Lab
or
0.7
00
.20
0.2
20
.65
0.2
80
.32
0.8
30
.38
0.4
20
.86
0.4
00
.44
0.8
90
.40
0.4
30
.78
0.3
30
.35
(0.0
4)
(0.0
1)
(0.0
2)
(0.0
6)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
2)
(0.0
2)
(0.0
6)
(0.0
3)
(0.0
5)
(0.0
4)
(0.0
2)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
1)
Cap
ital0
.33
0.0
80
.12
0.3
60
.15
0.1
60
.16
0.0
70
.05
0.1
20
.05
0.0
40
.25
0.1
10
.10
0.3
10
.13
0.1
4
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
4)
(0.0
2)
(0.0
2)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
4)
(0.0
2)
(0.0
2)
(0.0
3)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
0)
(0.0
1)
Interm
ediates
----
0.6
7--
--0
.54
----
0.5
2--
--0
.51
----
0.5
3--
--0
.54
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
0)
Sum
1.0
31
.01
1.0
11
.01
1.0
01
.01
0.9
90
.99
0.9
90
.98
0.9
90
.99
1.1
41
.06
1.0
61
.09
1.0
41
.04
(0.0
3)
(0.0
1)
(0.0
1)
(0.0
4)
(0.0
2)
(0.0
2)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
7)
(0.0
3)
(0.0
4)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
0)
(0.0
0)
Mean
(Cap
ital) /
Mean
(Lab
or)
0.4
70
.42
0.5
50
.55
0.5
50
.49
0.1
90
.19
0.1
20
.14
0.1
40
.08
0.2
80
.27
0.2
30
.39
0.3
80
.40
(0.0
6)
(0.0
5)
(0.0
8)
(0.1
0)
(0.1
0)
(0.0
9)
(0.0
3)
(0.0
3)
(0.0
4)
(0.0
5)
(0.0
5)
(0.0
5)
(0.0
4)
(0.0
4)
(0.0
4)
(0.0
2)
(0.0
2)
(0.0
3)
Ch
ile
Lab
or
0.7
70
.21
0.2
80
.93
0.3
70
.45
0.9
50
.37
0.4
50
.92
0.3
20
.40
0.9
60
.43
0.5
20
.77
0.2
90
.38
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
4)
(0.0
2)
(0.0
3)
(0.0
4)
(0.0
2)
(0.0
2)
(0.0
4)
(0.0
1)
(0.0
2)
(0.0
4)
(0.0
2)
(0.0
3)
(0.0
1)
(0.0
1)
(0.0
1)
Cap
ital0
.33
0.1
00
.11
0.2
40
.10
0.1
10
.20
0.0
80
.06
0.1
90
.07
0.0
70
.25
0.1
10
.13
0.3
70
.14
0.1
6
(0.0
1)
(0.0
0)
(0.0
1)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
3)
(0.0
1)
(0.0
1)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
2)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
0)
(0.0
0)
Interm
ediates
----
0.6
7--
--0
.54
----
0.5
6--
--0
.59
----
0.5
0--
--0
.55
(0.0
0)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
1)
(0.0
0)
Sum
1.1
01
.03
1.0
51
.17
1.0
71
.10
1.1
41
.06
1.0
81
.11
1.0
41
.06
1.2
21
.09
1.1
51
.13
1.0
51
.09
(0.0
2)
(0.0
0)
(0.0
1)
(0.0
3)
(0.0
1)
(0.0
2)
(0.0
3)
(0.0
1)
(0.0
2)
(0.0
3)
(0.0
1)
(0.0
1)
(0.0
3)
(0.0
1)
(0.0
2)
(0.0
1)
(0.0
0)
(0.0
1)
Mean
(Cap
ital) /
Mean
(Lab
or)
0.4
30
.46
0.3
90
.26
0.2
60
.24
0.2
10
.21
0.1
40
.21
0.2
20
.18
0.2
60
.27
0.2
50
.48
0.4
90
.43
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
4)
(0.0
4)
(0.0
4)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
3)
(0.0
2)
(0.0
2)
(0.0
2)
No
tes:
Wood
Pro
du
cts
(331)
Fab
ricated
Meta
ls
(381)
All
d. T
he ro
w titled
"Sum
" reports th
e sum
of th
e averag
e labor, cap
ital, and interm
ediate inp
ut elasticities, and
the ro
w titled
"Mean(C
apital)/M
ean(Lab
or)" rep
orts th
e ratio o
f the av
erage cap
ital elasticity to
the av
erage lab
or elasticity
.
a. Stand
ard erro
rs are estimated
using
the b
ootstrap
with
200 rep
lications and
are reported
in parenth
eses belo
w th
e point estim
ates.
b. F
or each
industry
, the nu
mbers in th
e first colu
mn are b
ased o
n a valu
e-added
specificatio
n and are estim
ated u
sing a co
mplete p
oly
nom
ial series of d
egree 2
with
the m
ethod fro
m A
ckerb
erg, C
aves, and
Frazer (2
006). T
he nu
mbers in th
e second
colu
mn are o
btained
by raising
the v
alue-ad
ded
estimates to
the p
ow
er of o
ne minu
s the firm
's share o
f intermed
iate inputs in to
tal outp
ut. T
he nu
mbers
in the th
ird co
lum
n are based
on a g
ross o
utp
ut sp
ecification and
are estimated
using
a com
plete p
oly
nom
ial series of d
egree 2
for each
of th
e two no
nparam
etric functio
ns (G and
C) o
f our ap
pro
ach.
c. Since th
e input elasticities are h
eterogeneo
us acro
ss firms, w
e report th
e averag
e input elasticities w
ithin each
giv
en industry
.
Tab
le 5: A
verag
e Input E
lasticities of O
utp
ut--R
escaled V
alue A
dded
(Stru
ctural E
stimates: R
escaled V
alue A
dded
vs. G
ross O
uput)
Industry
(ISIC
Co
de)
Food
Pro
du
cts
(311)
Tex
tiles
(321)
Ap
parel
(322)
55
Co
lom
bia
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
Valu
e
Added
(AC
F)
Valu
e
Added
(AC
F--
Rescaled
)
Gro
ss
Outp
ut
(GN
R)
75/2
5 ratio
2.2
02.2
91.3
31.9
71.8
51.3
51.6
61.8
81.2
91.7
32.7
41.3
01.7
82.0
01.3
11.9
52.1
51.3
7(0
.07
)(0
.12
)(0
.02
)(0
.09
)(0
.12
)(0
.03
)(0
.03
)(0
.09
)(0
.01
)(0
.08
)(0
.44
)(0
.04
)(0
.04
)(0
.10
)(0
.02
)(0
.17
)(0
.06
)(0
.01
)
90/1
0 ratio
5.1
74.9
91.7
73.7
13.2
81.8
32.8
73.6
01.6
63.0
86.2
71.8
03.3
33.8
11.7
44.0
14.6
81.8
6(0
.27
)(0
.51
)(0
.05
)(0
.30
)(0
.39
)(0
.07
)(0
.09
)(0
.34
)(0
.03
)(0
.38
)(1
.51
)(0
.12
)(0
.13
)(0
.35
)(0
.03
)(0
.07
)(0
.24
)(0
.02
)
95/5
ratio11.0
18.7
92.2
46.3
65.0
02.3
84.3
65.4
12.0
24.5
810.5
32.2
45.3
15.8
52.1
66.8
67.9
12.3
6(1
.11
)(1
.20
)(0
.08
)(0
.76
)(0
.87
)(0
.14
)(0
.22
)(0
.67
)(0
.05
)(1
.01
)(3
.19
)(0
.22
)(0
.34
)(0
.69
)(0
.06
)(0
.02
)(0
.52
)(0
.03
)
Expo
rter3.6
20.8
40.1
40.2
00.0
70.0
20.1
6-0
.04
0.0
50.2
60.5
50.1
50.2
00.0
60.0
80.5
10.0
90.0
6(0
.99
)(0
.30
)(0
.05
)(0
.10
)(0
.08
)(0
.03
)(0
.07
)(0
.07
)(0
.03
)(0
.63
)(0
.60
)(0
.14
)(0
.05
)(0
.07
)(0
.03
)(0
.12
)(0
.06
)(0
.01
)
Impo
rter-0
.25
-0.4
00.0
40.2
70.0
80.0
50.2
9-0
.09
0.1
20.0
6-0
.20
0.0
50.2
60.1
50.1
00.2
00.0
50.1
1(0
.08
)(0
.08
)(0
.02
)(0
.10
)(0
.08
)(0
.04
)(0
.08
)(0
.05
)(0
.03
)(0
.53
)(0
.20
)(0
.08
)(0
.06
)(0
.04
)(0
.02
)(0
.05
)(0
.04
)(0
.01
)
Advertiser
-0.4
6-0
.42
-0.0
30.2
0-0
.16
0.0
80.1
3-0
.28
0.0
50.0
2-0
.24
0.0
40.1
5-0
.05
0.0
5-0
.13
-0.1
80.0
3(0
.10
)(0
.10
)(0
.02
)(0
.07
)(0
.07
)(0
.03
)(0
.04
)(0
.06
)(0
.02
)(0
.09
)(0
.13
)(0
.04
)(0
.04
)(0
.04
)(0
.02
)(0
.06
)(0
.03
)(0
.01
)
Wag
es > M
edian
0.5
90.1
50.0
90.6
00.3
50.1
80.4
10.3
10.1
80.3
40.1
90.1
50.5
50.2
70.2
20.6
30.2
90.2
0(0
.19
)(0
.18
)(0
.02
)(0
.09
)(0
.08
)(0
.03
)(0
.03
)(0
.05
)(0
.02
)(0
.17
)(0
.18
)(0
.04
)(0
.06
)(0
.04
)(0
.02
)(0
.05
)(0
.04
)(0
.01
)
Ch
ile
75/2
5 ratio
2.9
22.3
61.3
72.5
62.0
61.4
82.5
82.9
41.4
33.0
63.3
11.5
02.4
52.4
71.5
33.0
02.9
21.5
5(0
.05
)(0
.11
)(0
.01
)(0
.07
)(0
.17
)(0
.02
)(0
.07
)(0
.27
)(0
.02
)(0
.08
)(0
.36
)(0
.02
)(0
.06
)(0
.19
)(0
.02
)(0
.03
)(0
.10
)(0
.01
)
90/1
0 ratio
9.0
25.2
41.9
06.7
73.8
02.1
66.7
68.0
82.1
110.1
210.4
72.3
26.2
75.7
22.3
39.1
97.2
42.3
9(0
.30
)(0
.44
)(0
.02
)(0
.30
)(0
.57
)(0
.05
)(0
.33
)(1
.44
)(0
.05
)(0
.60
)(2
.23
)(0
.05
)(0
.27
)(0
.85
)(0
.05
)(0
.15
)(0
.44
)(0
.02
)
95/5
ratio21.2
98.9
12.4
813.5
65.5
42.9
114.2
113.8
92.7
725.0
819.8
33.1
112.5
29.4
03.1
320.9
012.2
33.3
1(0
.99
)(0
.96
)(0
.05
)(0
.84
)(1
.05
)(0
.09
)(0
.77
)(2
.94
)(0
.09
)(2
.05
)(5
.41
)(0
.11
)(0
.78
)(1
.84
)(0
.10
)(0
.47
)(0
.91
)(0
.04
)
Expo
rter0.2
70.3
40.0
20.0
7-0
.03
0.0
20.1
80.1
40.0
90.1
20.0
90.0
00.0
30.0
1-0
.01
0.2
00.1
20.0
3(0
.10
)(0
.12
)(0
.02
)(0
.07
)(0
.06
)(0
.03
)(0
.08
)(0
.11
)(0
.03
)(0
.12
)(0
.11
)(0
.03
)(0
.06
)(0
.07
)(0
.03
)(0
.04
)(0
.05
)(0
.01
)
Impo
rter0.7
10.4
40.1
40.2
20.1
00.1
00.3
10.1
80.1
40.4
40.2
10.1
50.3
00.1
90.1
10.4
60.3
70.1
5(0
.11
)(0
.15
)(0
.02
)(0
.05
)(0
.05
)(0
.02
)(0
.05
)(0
.07
)(0
.02
)(0
.10
)(0
.12
)(0
.03
)(0
.05
)(0
.07
)(0
.02
)(0
.03
)(0
.04
)(0
.01
)
Advertiser
0.1
80.0
50.0
40.0
90.0
00.0
40.1
50.0
60.0
60.0
4-0
.04
0.0
30.0
70.0
30.0
10.1
40.1
40.0
6(0
.05
)(0
.05
)(0
.01
)(0
.04
)(0
.04
)(0
.02
)(0
.04
)(0
.06
)(0
.02
)(0
.04
)(0
.06
)(0
.01
)(0
.04
)(0
.06
)(0
.02
)(0
.02
)(0
.02
)(0
.01
)
Wag
es > M
edian
1.2
30.6
60.2
10.4
70.3
40.1
90.6
20.5
70.2
20.6
80.5
30.2
10.5
60.4
70.2
20.9
90.8
90.3
0(0
.09
)(0
.07
)(0
.01
)(0
.06
)(0
.06
)(0
.02
)(0
.06
)(0
.09
)(0
.02
)(0
.08
)(0
.11
)(0
.02
)(0
.06
)(0
.09
)(0
.02
)(0
.04
)(0
.05
)(0
.01
)
Notes:
b. F
or each
indu
stry, th
e nu
mbers in
the first co
lum
n are b
ased o
n a v
alue-ad
ded
specificatio
n an
d are estim
ated u
sing a co
mplete p
oly
nom
ial series of d
egree 2
with
the m
ethod fro
m A
ckerb
erg, C
aves, an
d F
razer (20
06
). Th
e nu
mbers in
the seco
nd co
lum
n are o
btain
ed b
y raisin
g th
e valu
e-added
estimates to
the p
ow
er of o
ne m
inu
s the firm
's share o
f interm
ediate in
pu
ts in to
tal
ou
tpu
t. Th
e nu
mbers in
the th
ird co
lum
n are b
ased o
n a g
ross o
utp
ut sp
ecification
and are estim
ated u
sing a co
mplete p
oly
nom
ial series of d
egree 2
for each
of th
e two n
on
param
etric fun
ction
s (G an
d C
) of o
ur ap
pro
ach.
c. In th
e first three ro
ws w
e report ratio
s of p
rodu
ctivity
for p
lants at v
ariou
s percen
tiles of th
e pro
du
ctivity
distrib
utio
n. In
the rem
ainin
g fo
ur ro
ws w
e report estim
ates of th
e pro
du
ctivity
differen
ces betw
een p
lants (as a fractio
n) b
ased o
n w
heth
er they
hav
e exported
som
e of th
eir ou
tpu
t, imported
interm
ediate in
pu
ts, spen
t mon
ey o
n ad
vertisin
g, an
d p
aid w
ages ab
ove th
e indu
stry
med
ian. F
or ex
ample, in
indu
stry 3
11
for C
hile v
alue ad
ded
implies th
at a firm th
at advertises is, o
n av
erage, 1
8%
more p
rodu
ctive th
an a firm
that d
oes n
ot ad
vertise.
Ap
parel
(322)
Wood
Pro
du
cts
(331)
Fab
ricated
Meta
ls
(381)
All
Tab
le 6: H
eterogen
eity in
Pro
du
ctivity
--Rescaled
Valu
e Ad
ded
(Stru
ctural E
stimates: R
escaled V
alue A
dd
ed v
s. Gro
ss Oup
ut)
Industry
(ISIC
Co
de)
Food
Pro
du
cts
(311)
Tex
tiles
(321)
a. Stan
dard
errors are estim
ated u
sing th
e bootstrap
with
20
0 rep
lication
s and are rep
orted
in p
arenth
eses belo
w th
e poin
t estimates.
56
Appendix A: The Olley-Pakes Estimator
The original Olley and Pakes (1996) approach differs from LP in that it aims to exploit the structure
of a predetermined input demand as an alternative to the flexible intermediate input as the proxy
variable. In particular they treat kt as the predetermined input in the model whereas the remaining
inputs (lt,mt in this case) are treated as flexible. By using demand for future capital (investment)
as a proxy,38 rather than intermediate inputs, it may appear that the identification problems we
raise above are less severe for the OP empirical strategy, as kt+1 is not a direct input into the
production function at t. However, we now show that capital as a proxy faces the same fundamental
identification problems we raise above.39
The key idea behind the OP approach is to assume that the structure of the predetermined input
demand satisfies the following scalar unobservability assumption
Assumption 6. The predetermined input demand is given by
kt = K (It−1) = K (kt−1, ωt−1)
where K is strictly monotone increasing in the second argument.
This assumption allows them to exploit kt+1 as a proxy variable for productivity ωt. This can
be seen by examining the “first stage” of their procedure in which they regress output yt on the
inputs (kt, lt,mt) as well as the proxy variable kt+1, which yields
E [yt | kt, lt,mt, kt+1] = f (kt,, lt,mt) + E [ωt | kt,, lt,mt, kt+1]
= f (kt,, lt,mt) + K−1 (kt+1, kt) , (32)
where the second equality follows from Assumption 6.
As Ackerberg, Caves, and Frazer (2015) have shown, since mt = M (kt, ωt) = M (kt, kt+1),
38Technically OP used investment it as the proxy, but given the process they assume for capital: Kt+1 =(1− depreciation)Kt + Ii, and conditional on kt, using kt+1 is equivalent to using it.
39This leaves aside altogether the original LP motivation that investment is often zero in the data and hence thoseobservations cannot be used for the estimation.
A-1
and similarly for labor, in general the OP procedure faces the same identification problems regard-
ing the flexible input elasticities. However, in contrast to LP, ACF show that the economic structure
on the input decisions in OP admits some model-consistent sources of variation that would permit
the elasticity ∂∂xtE [yt | kt,, lt,mt, kt+1] to be identified in the data for xt ∈ {lt,mt}. If such sources
of variation exist, then for xt ∈ {lt,mt} the elasticity
∂
∂xtE [yt | kt,, lt,mt, kt+1] =
∂
∂xtf (kt,, lt,mt) (33)
is identified in the data. Notice also that since K−1 (kt+1, kt) = ωt in equation (32), it follows that
the ex-post productivity shock, εt = yt − E [yt | kt,, lt,mt, kt+1], is also identified.
The capital elasticity clearly cannot be identified from the first stage (33). The empirical strat-
egy proposed by OP for the capital elasticity makes use of parametric structure of the production
function, which they assumed took a Cobb-Douglas form (C-D for short):
f (kt,, lt,mt) = αkkt + αllt + αmmt.
The first stage of the model in this case is
E [yt | kt,, lt,mt, kt+1] = αkkt + αllt + αmmt + K−1 (kt+1, kt)
= αllt + αmmt + φ (kt+1, kt) (34)
which takes the form of a partially linear model (Robinson, 1988). The random variable φt ≡
φ (kt+1, kt) can be recovered in the data from the partially linear regression in (34), as can its
derivatives. The “second stage” of the OP empirical strategy to identify the capital elasticity is
based on regressing the new “dependent variable”
yt ≡ yt − αllt − αmmt
A-2
on (kt, φt−1, kt−1). This second stage thus yields
E [y | kt, kt−1, φt−1] = αkkt + E [ωt | kt, kt−1, φt−1]
= αkkt + h (φt−1 − αkkt−1) . (35)
As we now show, however, under the model’s restrictions, the joint variation in (kt, kt−1, φt−1)
required to identify the capital elasticity αk is unlikely to exist in the data. The reason is based on
the following observation. Because ωt−1 = φt−1 − αkkt−1, which implies
kt = K (kt−1, ωt−1) = K (kt−1, φt−1) , (36)
there is no variation in kt conditional on (kt−1, φt−1).40 Therefore, the partial derivative
∂
∂ktE [y | kt, kt−1, φt−1] = αk (37)
cannot be recovered directly in the data.
This observation gives rise to two fundamental problems for the OP solution to transmission
bias. The first problem is that any real data application is unlikely to actually exhibit the knife-edge
variation (36) implied by the model, i.e., real data will typically exhibit variation in kt conditional
on (kt−1, φt−1), in which case one could recover the LHS of (37). Applying their estimator to such
data would estimate αk using variation that the model predicts should not exist. The estimates of
“αk” in this case will have no structural meaning (i.e., will not correspond to the capital elasticity)
as the data rejects the model.
This identification problem regarding the OP approach can be seemingly alleviated if we try to
impose a version of Assumption 4 to generate an additional source of variation in kt conditional on
(kt−1, φt−1). If the model allowed for another source of variation χt that enters the predetermined
40This is in essence what Assumption 4 rules out in the nonparametric case.
A-3
input demand function, we would have
kt = K (It−1) = K (kt−1, ωt−1, χt−1) .
The second problem is that such variation χt would represent additional dimensions of firm hetero-
geneity in addition to productivity ωt, and would thus violate the scalar unobservability assumption
that drives the OP approach.
If χt were observed and augmented to the data, then using it to resolve the identification prob-
lem would require that χt accounts for all the variation in kt conditional on kt−1 and φt−1. Any
unexplained variation would again violate the scalar unobservability assumption on the prede-
termined input. Even if this requirement is met, χt will likely be a higher dimensional random
variable that must now also be included in the nonparametric estimation of φ (kt, kt+1, χt). This
places a major burden on the data to estimate a high-dimensional nonparametric function. This is
especially problematic in applications because, as pointed out by LP, real data sets often contain a
large percentage of observations with zero investment which must be excluded from the empirical
estimation of φt, thus creating a severe small sample problem that would undermine this strategy.
Appendix B: A Parametric Example
In order to illustrate our non-identification result, we consider a parametric example. Suppose
that the true production function is Cobb-Douglas F (kt, lt,mt) = Kαkt Lαlt M
αmt , and productivity
follows an AR(1) process ωt = δ0 + δωt−1 + ηt.41 The parametric version of the instrumental
variables restriction in equation (6) is the following:
E [yt | Γt] = constant+αkkt+αllt+αmE [mt | Γt]+δ0+δ (φt−1 − αkkt−1 − αllt−1 − αmmt−1) ,
41For simplicity we assume that the prices of output and intermediate inputs are non-time-varying. While time-series variation in relative prices would provide a source of identifying variation here, as we discuss in footnote 21,relying on only time-series variation in prices is problematic for several reasons.
A-4
where φt−1 = f (kt−1, lt−1,mt−1) + M−1 (kt−1, lt−1,mt−1). If we plug in for mt using the first-
order condition and combine constants we have
E [yt | Γt] = ˜constant+ αkkt + αllt + αm
(αkkt + αllt + δ (φt−1 − f (kt−1, lt−1,mt−1))
1− αm
)+δ (φt−1 − αkkt−1 − αllt−1 − αmmt−1)
= ˜constant+
(αk
1− αm
)kt +
(αl
1− αm
)lt
+δ
1− αm(φt−1 − αkkt−1 − αllt−1 − αmmt−1) .
Plugging in for the Cobb-Douglas parametric form of M−1, it can be shown that φt−1 = mt−1,
which implies
E [yt | Γt] = ˜constant+
(αk
1− αm
)kt+
(αl
1− αm
)lt−δ
(αk
1− αm
)kt−1−δ
(αl
1− αm
)lt−1−δmt−1.
Notice that, although there are five sources of variation (kt, lt, kt−1, lt−1,mt−1), the model is
not identified. Variation in mt−1 identifies δ, but the coefficient on kt is equal to the coefficient on
kt−1 multiplied by −δ, and the same is true for l. In other words, variation in kt−1 and lt−1 do not
provide any additional information about the parameters of the production function. As a result, all
we can identify is αk(
11−αm
)and αl
(1
1−αm
). To put it another way, the rank condition necessary
for identification of this model is not satisfied.
In terms of our proposed alternative structure in Theorem 1, we would have
αk = (1− a)αk ; αl = (1− a)αl ; αm = (1− a)αm + a ; δ = δ .
It immediately follows that(
αk1−αm
)=(
αk1−αm
)and
(αl
1−αm
)=(
αl1−αm
), and thus our continuum
of alternative structures indexed by a ∈ (0, 1) satisfy the instrumental variables restriction.
Doraszelski and Jaumandreu (2013) avoid this problem by exploiting both parametric restric-
tions and observed price variation as an instrument for identification. This illustrates an important
difference with our approach. In addition to not requiring parametric restrictions (or price varia-
A-5
tion), we are not using the first-order condition to find a replacement function for ω in the produc-
tion function. Instead, we use it to form the share regression equation, which gives us a second
structural equation that we use in identification and estimation. In terms of our Cobb-Douglas
example, the second equation would be given by the following share equation st = αm− εt. Since
this equation identifies αm (given that E [εt] = 0), this is enough to allow for identification of the
whole production function and productivity.
Appendix C: Extensions
In this section we discuss four extensions to our baseline model: allowing for fixed effects, in-
corporating additional unobservables in the flexible input demand, allowing for multiple flexible
inputs, and revenue production functions.
C1. Fixed Effects
One benefit of our identification strategy is that it can easily incorporate fixed effects in the pro-
duction function. With fixed effects, the production function can be written as
yt = f (kt, lt,mt) + a+ ωt + εt, (38)
where a is a firm-level fixed effect.42 From the firm’s perspective, the optimal decision problem for
intermediate inputs is the same as before, as is the derivation of the nonparametric share regression
(equation 15), with ωt ≡ a+ ωt replacing ωt.
The other half of our approach can be easily augmented to allow for the fixed effects. We
follow the dynamic panel data literature and impose that persistent productivity ω follows a first-
order linear Markov process to difference out the fixed effects: ωt = δωt−1 + ηt.43 The equivalent
42Kasahara, Schrimpf, and Suzuki (2015) generalize our approach to allow for the entire production function to befirm-specific.
43For simplicity we use an AR(1) here, but higher order auto-regressive models can be incorporated as well. Weomit the constant from the Markov process since it is not separately identified from the mean of the fixed effects.
A-6
of equation (9) is given by:
Yt = a− C (kt, lt) + δ (Yt−1 + C (kt−1, lt−1)) + ηt.
Subtracting the counterpart for period t − 1 eliminates the fixed effect. Re-arranging terms leads
to:
Yt − Yt−1 = − (C (kt, lt)− C (kt−1, lt−1)) + δ (Yt−1 − Yt−2)
+δ (C (kt−1, lt−1)− C (kt−2, lt−2)) + (ηt − ηt−1) .
Recall that E [ηt | Γt] = 0. Since Γt−1 ⊂ Γt, this implies that E [ηt − ηt−1 | Γt−1] = 0, where Γt−1
includes (kt−1, lt−1,Yt−2, kt−2, lt−2,Yt−3, ...).
Let
r (kt, lt, kt−1, lt−1, (Yt−1 − Yt−2) , kt−2, lt−2) = − (C (kt, lt)− C (kt−1, lt−1)) (39)
+δ (Yt−1 − Yt−2)
+δ (C (kt−1, lt−1)− C (kt−2, lt−2)) .
From this we have the following nonparametric IV equation
E [Yt − Yt−1 | kt−1, lt−1,Yt−2, kt−2, lt−2, kt−3, lt−3]
= E [r (kt, lt, kt−1, lt−1, (Yt−1 − Yt−2) , kt−2, lt−2) | kt−1, lt−1,Yt−2, kt−2, lt−2, kt−3, lt−3] ,
which is an analogue to equation (11), in the case without fixed effects.
Theorem 5. Under Assumptions 2 - 5, plus the additional assumption that the distribution of the
endogenous variables conditional on the exogenous variables (i.e., instruments),
G (lt, kt, lt−1, kt−1, (Yt−1 − Yt−2) , lt−2, kt−2 | lt−3, kt−3, lt−1, kt−1,Yt−2, lt−2, kt−2), is complete (as
defined in Newey and Powell, 2003), the production function f is nonparametrically identified up
to an additive constant if ∂∂mt
f (lt, kt,mt) is nonparametrically known.
A-7
Following the first part of the proof of Theorem 3, we know that the production function is
identified up to an additive function C (kt, lt). Following directly from Newey and Powell (2003),
we know that, if the distribution G is complete, then the function r defined in equation (39) is
identified.
Let(C , δ
)be a candidate alternative structure. The two structures (C , δ) and
(C , δ
)are
observationally equivalent if and only if
− (C (kt, lt)− C (kt−1, lt−1)) + δ (Yt−1 − Yt−2) + δ (C (kt−1, lt−1)− C (kt−2, lt−2))
= −(C (kt, lt)− C (kt−1, lt−1)
)+ δ (Yt−1 − Yt−2) + δ
(C (kt−1, lt−1)− C (kt−2, lt−2)
).
(40)
By taking partial derivatives of both sides of (40) with respect to kt and lt we obtain
∂
∂zC (kt, lt) =
∂
∂zC (kt, lt)
for z ∈ {kt, lt}, which implies C (kt, lt)− C (kt, lt) = c for a constant c. Thus we have shown the
production function is identified up to an additive constant.
The estimation strategy for the model with fixed effects is almost exactly the same as without
fixed effects. The first stage, estimating Dr (kt, lt,mt), is the same. We then form Yt in the same
way. We also use the same series estimator for C (kt, lt). This generates analogues to equations
(22) and (23):
Yt−Yt−1 = −∑
0<τk+τl≤τ
ατk,τlkτkt l
τlt +
∑0<τk+τl≤τ
ατk,τlkτkt−1l
τlt−1+δ (ωt−1 − ωt−2)+(ηt − ηt−1) (41)
andYt − Yt−1 = −
∑0<τk+τl≤τ ατk,τlk
τkt l
τlt + δ (Yt−1 − Yt−2)
+ (δ + 1)(∑
0<τk+τl≤τ ατk,τlkτkt−1l
τlt−1
)−δ(∑
0<τk+τl≤τ ατk,τlkτkt−2l
τlt−2
)+ (ηt − ηt−1) .
(42)
We can use similar moments as for the model without fixed effects, except that now we need to lag
the instruments one period given the differencing involved. Therefore the following moments can
A-8
be used to form a standard sieve GMM criterion function to estimate (α, δ): E [(ηt − ηt−1) kτkt−ιlτlt−ι],
for ι ≥ 1.
C2. Extra Unobservables
Our identification and estimation approach can also be extended to incorporate additional unob-
servables driving the intermediate input demand.
In our baseline model, our system of equations consists of the share equation and the production
function given by
st = lnD (kt, lt,mt) + ln E − εt
yt = f (kt, lt,mt) + ωt + εt.
We now show that our model can be extended to include an additional structural unobservable to
the share equation for intermediate inputs, which we denote by ψt:
st = lnD (kt, lt,mt) + ln E − εt − ψt (43)
yt = f (kt, lt,mt) + ωt + εt,
where E ≡ E[eψt+εt
].
Assumption 7. ψt ∈ It is known to the firm at the time of making its period t decisions and is not
persistent: Pψ (ψt | It−1) = Pψ (ψt).
C2.1. Interpretations for the extra unobservable
We now discuss some possible interpretations for the extra unobservable ψ.
Shocks to prices of output and/or intermediate inputs Suppose that the prices of output and
intermediate inputs, Pt and ρt, are not fully known when firm j decides its level of intermediate
A-9
inputs, but that the firm has private signals about the prices, denoted P ∗jt and ρ∗jt, where
lnP ∗jt = lnPt − νjt,
ln ρ∗jt = ln ρt − νMjt ,
where we add the firm subscripts j for clarity. Firms maximize expected profits conditional on
their signals:
M (kjt, ljt, ωt) = maxMjt
Eε,ν,νM[PtF (kjt, ljt,mjt) e
ωjt+εjt − ρtMjt | P ∗jt, ρ∗jt]
= maxMjt
Eε,ν,νM[(P ∗jte
νjt)F (kjt, ljt,mjt) e
ωjt+εjt −(ρ∗jte
νMjt
)Mjt | P ∗jt, ρ∗jt
]= max
Mjt
E (eνjt)E (eεjt)P ∗jtF (kjt, ljt,mjt) eωjt − E
(eν
Mjt
)ρ∗jtMjt.
This implies that the firm’s first order condition for intermediate inputs is given by
E (eνjt)E (eεjt)P ∗jt∂
∂Mjt
F (kjt, ljt,mjt) eωjt − E
(eν
Mjt
)ρ∗jt = 0,
which can be rewritten as
(ρtMjt
PtYjt
)=E (eνjt)E (eεjt)
E(eν
Mjt
) ∂
∂mjt
f (kjt, ljt,mjt)eν
Mjt
evjteεjt.
Let ψjt ≡ νjt − νMjt , and we have
lnρtMjt
PtYjt= sjt = lnD (kjt, ljt,mjt) + ln E − εjt − ψjt.
Optimization error Suppose that firms do not exactly know their productivity, ωt, when they
make their intermediate input decision. Instead, they observe a signal about productivity ω∗t =
ωt − ψt, where ψt denotes the noise in the signal. The firm’s profit maximization problem with
A-10
respect to intermediate inputs is
M (kt, lt, ωt) = arg maxMt
PtEε,ψ[F (kt, lt,mt) e
ω∗t+ψt+εt]− ρtMt.
This implies the following first order condition
Pt∂
∂Mt
F (kt, lt,mt) eω∗tEε,ψ
[eψt+εt
]= ρt
Re-arranging to solve for the share of intermediate inputs gives us the share equation
lnρtMt
PtYt= st = lnD (kt, lt,mt) + ln E − εt − ψt.
Notice that for both interpretations of ψ, the firm will take into account the value of E[eεt+ψt
]when deciding on the level of intermediate inputs, which means we want to correct the share
estimates by this term. As in the baseline model, we can recover this term by estimating the share
equation, forming the residuals, εt + ψt, and computing the expectation of eεt+ψt .
C2.2. Identification
The identification of the share equation is similar to our main specification, but with two differ-
ences. The first is that, since ψt drives intermediate input decisions and is in the residual of the
modified share equation (43), intermediate inputs are now endogenous in the share equation. As
a result, we need to instrument for mt in the share regression. We can use mt−1 as an instrument
for mt, since it is correlated with mt and independent of the error (εt + ψt). Since in the share re-
gression we condition only on kt and lt (and no lags), mt−1 generates variation in mt (conditional
on kt and lt), due to Assumptions 3 and 4. Identification follows from standard nonparametric IV
arguments as in Newey and Powell (2003).
The second difference is that the error in the share equation is εt + ψt instead of εt. We can
A-11
form an alternative version of Yt, which we denote Yt:
Yt ≡ yt −∫D (kt, lt,mt) dmt − (εt + ψt) = Yt − ψt. (44)
This generates an analogous equation to equation (8) in the paper:
Yt = −C (kt, lt) + ωt − ψt ⇒ ωt = Yt + C (kt, lt) + ψt.
Re-arranging terms and plugging in the Markovian structure of ω gives us:
Yt = −C (kt, lt) + h
Yt−1 + C (kt−1, lt−1) + ψt−1︸ ︷︷ ︸ωt−1
+ ηt − ψt, (45)
which is an analogue of equation (9).
The challenge is that we cannot form ωt−1, the argument of h in equation (45), because ψt−1
is not observed. We can, however, construct two noisy measures of ωt−1: (ωt−1 + εt−1) and
(ωt−1 − ψt−1) where
ωt−1 + εt−1 = yt−1 − f (kt−1, lt−1,mt−1)
= yt−1 −∫D (kt−1, lt−1,mt−1) dmt−1 + C (kt−1, lt−1)
ωt−1 − ψt−1 = (ωt−1 + εt−1)− (εt−1 + ψt−1)
=
(yt−1 −
∫D (kt−1, lt−1,mt−1) dmt−1 + C (kt−1, lt−1)
)− (st−1 − lnD (kt−1, lt−1,mt−1)) .
We could proceed to identify h and C from equation (45) by adopting methods from the mea-
surement error literature (Hu and Schennach (2008) and Cunha, Heckman, and Schennach (2010))
using one of the noisy measures as our measure of ωt−1 and using the other as an instrument.
However, such an exercise is beyond the scope of the current paper.
A-12
Instead, we illustrate our approach using an AR(1) process for the evolution of ω: h (ωt−1) =
δ0 + δωt−1 + ηt. We can then re-write equation (45) as
Yt = −C (kt, lt) + δ0 + δ(Yt−1 + C (kt−1, lt−1)
)+ ηt − ψt + δψt−1, (46)
where now the residual is given by ηt − ψt + δψt−1. Given Assumptions 2 and 7, we have that
E [ηt − ψt + δψt−1 | Γt−1] = 0, where recall that Γt−1 = Γ (It−2), i.e., a transformation of the
period t− 2 information set. If we let
r(kt, lt, Yt−1, kt−1, lt−1
)= −C (kt, lt) + δ0 + δ
(Yt−1 + C (kt−1, lt−1)
),
then identification of equation (46) follows from a parallel argument to that in Theorem 5 (i.e., in-
cluding the completeness assumption and following the nonparametric IV identification arguments
in Newey and Powell (2003)). Therefore we can identify the entire production function, up to an
additive constant. We can also identify δ0 and δ, as well as productivity: ω + ε.
C3. Multiple Flexible Inputs
Suppose that, in addition to intermediate inputs being flexible, the researcher believes that one
or more additional inputs are also flexible.44 Our approach can also be extended to handle this
case. In what follows we assume that labor is the additional flexible input, but the approach can be
extended to allow for more than two flexible inputs.
When labor and intermediate inputs are both assumed to be flexible, we have two share equa-
tions. We use superscripts M and L to distinguish them. Given the extra equation, we allow for
additional structural errors in the model, ψ, as described in the preceding sub-section. Our system
44See, for example, Doraszelski and Jaumandreu (2013).
A-13
of equations is thus given by:
sMt = lnDM (kt, lt,mt) + ln EM − εt − ψMt
sLt = lnDL (kt, lt,mt) + ln EL − εt − ψLt
yt = f (kt, lt,mt) + ωt + εt.
Nonparametric identification of the flexible input elasticities of L and M proceeds as above in
Appendix C2.
These two input elasticities define a system of partial differential equations of the production
function. By the fundamental theorem of calculus we have
∫ mt
m0
∂
∂mt
f (kt, lt,mt) dmt = f (kt, lt,mt) + CM (kt, lt)
and
∫ lt
l0
∂
∂ltf (kt, lt,mt) dlt = f (kt, lt,mt) + C L (kt,mt)
where now we have two constants of integration, one for each integrated share equation, CM (kt, lt)
and C L (kt,mt). Following directly from Varian (1992), these partial differential equations can be
combined to construct the production function as follows:
f (kt, lt,mt) =
mt∫m0
∂
∂mt
f (kt, l0, s) ds+
lt∫l0
∂
∂ltf (kt, τ,mt) dτ − C (kt) . (47)
That is, by integrating the (log) elasticities of intermediate inputs and labor, we can construct the
production function up to a constant that is a function of capital only.45
45In order to see why this is the case, evaluate the integrals on the RHS of equation (47), we have the following
f (kt, lt,mt) =(f (kt, l0,mt)− CM (kt, l0)
)−(f (kt, l0,m0)− CM (kt, l0)
)+(f (kt, lt,mt)− C L (kt,mt)
)−(f (kt, l0,mt)− C L (kt,mt)
)+f (kt, l0,m0)
= f (kt, lt,mt) ,
A-14
We can now construct an analogue to equation (44) above using the residual from either share
equation. Using the intermediate input share equation, we have
˜Yt ≡ yt −mt∫
m0
∂
∂mt
f (kt, l0, s) ds−lt∫
l0
∂
∂ltf (kt, τ,mt) dτ −
(εt + ψMt
)(48)
By subtracting equation (48) from the production function and re-arranging terms we have
˜Yt = −C (kt) + ωt − ψMt .
Plugging in the Markovian structure of ω gives us
˜Yt = −C (kt) + h
˜Y t−1 + C (kt−1) + ψMt−1︸ ︷︷ ︸ωt−1
+ ηt − ψMt , (49)
an analogue to equation (45) above. Identification of C and h can be achieved in the same way as
described in C2 for equation (45), with the difference that in this case C only depends on capital.
C4. Revenue Production Functions
We now show that our empirical strategy can be extended to the setting with imperfect competition
and revenue production functions such that 1) we solve the identification problem with flexible
inputs and 2) we can recover time-varying industry markups.46 We specify a generalized version
of the demand system in Klette and Griliches (1996) and De Loecker (2011),
PjtΠt
=
(YjtYt
) 1σt
eχjt , (50)
where f (kt, l0,m0) ≡ C (kt) is a constant of integration that is a function of capital kt.46This stands in contrast to the Klette and Griliches (1996) approach that can only allow for a markup that is time-
invariant.
A-15
where Pjt is the output price of firm j, Πt is the industry price index, Yt is a quantity index that
plays the role of an aggregate demand shifter,47 χjt is an observable (to the firm) demand shock,
and σt is the elasticity of demand that is allowed to vary over time.
Substituting for price using equation (50), the firm’s first order condition with respect to Mjt
in the (expected) profit maximization problem is
(1
σt+ 1
)Πt
Y1σtjt
Y1σtt
1
e1σtεjt
∂
∂Mjt
F (kjt, ljt,mjt) eχjtE
[eεjt
(1σt
+1)]
= ρt.
Following the same strategy as before, we can rewrite this expression in terms of the observed log
revenue share, which becomes
sjt = ln
(1
σt+ 1
)+ ln
(D (kjt, ljt,mjt)E
[eεjt
(1σt
+1)])−(
1
σt+ 1
)εjt, (51)
where sjt ≡ ln(ρtMt
PjtYt
), 1(
1σt
+1) is the expected markup, D (·) is the output elasticity of interme-
diate inputs, and εjt is the ex-post shock. Equation (51) nests the one obtained for the perfectly
competitive case in (15), the only difference being the addition of the expected markup, which is
equal to 1 under perfect competition.
We now show how to use the share regression (51) to identify production functions among
imperfectly competitive firms. Letting εjt =(
1σt
+ 1)εjt, equation (51) becomes
sjt = Υt + lnD (kjt, ljt,mjt) + ln E − εjt, (52)
where E = E[eεjt]
and Υt = ln(
1σt
+ 1)
. The intermediate input elasticity can be rewritten so
that we can break it into two parts: a component that varies with inputs and a constant µ, i.e.,
47As noted by Klette and Griliches (1996) and De Loecker (2011), Yt can be calculated using a market-shareweighted average of deflated revenues.
A-16
lnD (kjt, ljt,mjt) = lnDµ (kjt, ljt,mjt) + µ. Then, equation (52) becomes
sjt = (Υt + µ) + ln E + lnDµ (kjt, ljt,mjt)− εjt
= ϕt + ln E + lnDµ (kjt, ljt,mjt)− εjt.(53)
As equation (53) makes clear, without observing prices, we can nonparametrically recover the
scaled ex-post shock εjt (and hence E), the output elasticity of intermediate inputs up to a constant
lnDµ(kjt, ljt,mjt) = lnD(kjt, ljt,mjt)−µ, and the time-varying markups up to the same constant,
ϕt = Υt + µ, using time dummies for ϕt. Recovering the growth pattern of markups over time is
useful as an independent result as it can, for example, be used to check whether market power has
increased over time, or to analyze the behavior of market power with respect to the business cycle.
As before, we can correct our estimates for E and solve the differential equation that arises
from equation (53). However, because we can still only identify the elasticity up to the constant µ,
we have to be careful about keeping track of it as we can only calculate∫Dµ (kjt, ljt,mjt) dmjt =
e−µ∫D (kjt, ljt,mjt) dmjt. It follows that
f (kjt, ljt,mjt) e−µ + C (kjt, ljt) e
−µ =
∫Dµ (kjt, ljt,mjt) dmjt.
From this equation it is immediately apparent that, without further information, we will not be able
to separate the integration constant C (kjt, ljt) from the unknown constant µ.
To see how both the constant µ and the constant of integration can be recovered, notice that
what we observe in the data is the firm’s real revenue, which in logs is given by rjt = (pjt − πt) +
yjt. Recalling equation (2), and replacing for pjt−πt using (50), the observed log-revenue produc-
tion function is
rjt =
(1
σt+ 1
)f(kjt, ljt,mjt)−
1
σtyt + χjt +
(1
σt+ 1
)ωjt + εjt. (54)
However, we can write(
1 + 1σt
)= eϕte−µ. We know ϕt from our analysis above, so only µ is
A-17
unknown. Replacing back into (54) we get
rjt = eϕte−µf (kjt, ljt,mjt)−(eϕte−µ − 1
)yt (55)
+[(eϕte−µ
)ωjt + χjt
]+ εjt.
We then follow a similar strategy as before. As in equation (8), we first form an observable
variable
Rjt ≡ ln
(PjtYjt
Πt
eεjteeϕt∫Dµ(kjt,ljt,mjt)dmjt
),
where we now use revenues (the measure of output we observe), include eϕt , as well as using Dµ
instead of the, for now, unobservable D. Replacing into (55) we obtain
Rjt = −eϕt−µC (kjt, ljt)−(eϕte−µ − 1
)yt +
[(eϕte−µ
)ωjt + χjt
].
From this equation it is clear that the constant µ will be identified from variation in the observed
demand shifter yt. Without having recovered ϕt from the share regression first, it would not be
possible to identify time-varying markups. Note that in equation (54) both σt and yt change with
time, and hence yt cannot be used to identify σt unless we restrict σt = σ as in Klette and Griliches
(1996) and De Loecker (2011).
Finally, we can only recover a linear combination of productivity and the demand shock,(1 + 1
σt
)ωjt + χjt. The reason is clear: since we do not observe prices, we have no way of
disentangling whether, after controlling for inputs, a firm has higher revenues because it is more
productive (ωjt) or because it can sell at a higher price (χjt). We can write ωµjt =(
1 + 1σt
)ωjt+χjt
as a function of the parts that remain to be recovered
ωµjt = Rjt + eϕt−µC (kjt, ljt) +(eϕte−µ − 1
)yt,
A-18
and impose the Markovian assumption on this combination:48 ωµjt = h(ωµjt−1
)+ ηµjt. We can use
similar moment restrictions as before, E(ηµjt|kjt, ljt
)= 0, to identify the constant of integration
C (kjt, ljt) as well as µ (and hence the level of the markups).
48In this case, one would need to replace Assumption 2 with the assumption that the weighted sum of productivityωjt and the demand shock, χjt is Markovian. Note that this assumption does not necessarily imply that the twocomponents will be Markovian individually. See De Loecker (2011) for an example that imposes this assumption.
A-19
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Labor 0.15 0.21 0.37 0.28 0.29 0.22
(0.02) (0.03) (0.03) (0.06) (0.03) (0.01)
Capital 0.06 0.05 0.05 0.01 0.04 0.08
(0.01) (0.02) (0.01) (0.03) (0.02) (0.01)
Raw Materials 0.71 0.69 0.49 0.63 0.58 0.63
(0.02) (0.03) (0.04) (0.07) (0.03) (0.01)
Energy+Services 0.08 0.11 0.09 0.10 0.11 0.11
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
Sum 1.01 1.05 1.00 1.02 1.02 1.04
(0.01) (0.02) (0.01) (0.03) (0.01) (0.01)
Mean(Capital) /
Mean(Labor) 0.43 0.23 0.12 0.04 0.14 0.37
(0.08) (0.14) (0.04) (0.08) (0.06) (0.04)
Chile
Labor 0.18 0.28 0.31 0.29 0.31 0.22
(0.02) (0.03) (0.03) (0.04) (0.02) (0.01)
Capital 0.06 0.08 0.04 0.06 0.08 0.11
(0.01) (0.01) (0.01) (0.02) (0.01) (0.01)
Raw Materials 0.77 0.65 0.65 0.59 0.63 0.67
(0.02) (0.02) (0.02) (0.05) (0.02) (0.01)
Energy+Services 0.07 0.07 0.06 0.11 0.07 0.07
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
Sum 1.08 1.08 1.06 1.05 1.10 1.08
(0.01) (0.01) (0.01) (0.02) (0.01) (0.00)
Mean(Capital) /
Mean(Labor) 0.36 0.28 0.13 0.20 0.26 0.49
(0.04) (0.05) (0.04) (0.06) (0.04) (0.03)
Notes:
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification in which energy+services is flexible and raw materials is quasi-fixed. The results are estimated using a complete polynomial series of
degree 2 for each of the two nonparametric functions (G and C ) of our approach.
c. Since the input elasticities are heterogeneous across firms, we report the average input elasticities within each given industry.
Table D1: Average Input Elasticities of Output--Energy+Services Flexible(Structural Estimates: Gross Ouput)
d. The row titled "Sum" reports the sum of the average labor, capital, raw materials, and energy+services elasticities, and the row titled "Mean(Capital)/Mean(Labor)" reports the ratio of the average capital
elasticity to the average labor elasticity.
Appendix D: Additional Results
A-20
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
75/25 ratio 1.20 1.25 1.24 1.30 1.28 1.33(0.02) (0.03) (0.03) (0.06) (0.02) (0.01)
90/10 ratio 1.50 1.62 1.60 1.75 1.68 1.83(0.05) (0.10) (0.07) (0.16) (0.06) (0.03)
95/5 ratio 1.87 2.09 2.00 2.26 2.04 2.43(0.09) (0.22) (0.11) (0.24) (0.11) (0.08)
Exporter 0.14 -0.04 0.02 0.14 0.08 0.01(0.04) (0.06) (0.03) (0.12) (0.02) (0.03)
Importer 0.00 -0.03 -0.03 -0.04 0.10 -0.02(0.02) (0.06) (0.03) (0.05) (0.02) (0.05)
Advertiser -0.12 -0.13 -0.10 -0.07 0.05 -0.16(0.03) (0.11) (0.04) (0.06) (0.02) (0.05)
Wages > Median 0.06 0.13 0.14 0.09 0.19 0.10(0.02) (0.05) (0.02) (0.05) (0.02) (0.04)
Chile
75/25 ratio 1.31 1.42 1.41 1.44 1.48 1.49(0.01) (0.02) (0.02) (0.04) (0.02) (0.01)
90/10 ratio 1.76 2.04 2.01 2.13 2.23 2.26(0.03) (0.05) (0.04) (0.12) (0.05) (0.02)
95/5 ratio 2.22 2.69 2.65 2.94 2.94 3.08(0.05) (0.12) (0.07) (0.21) (0.08) (0.04)
Exporter -0.01 -0.06 0.01 0.01 -0.05 -0.03(0.03) (0.03) (0.03) (0.05) (0.03) (0.01)
Importer 0.02 0.04 0.07 0.09 0.05 0.07(0.02) (0.02) (0.02) (0.04) (0.02) (0.01)
Advertiser -0.01 -0.01 0.01 0.01 0.00 0.02(0.01) (0.02) (0.01) (0.01) (0.02) (0.01)
Wages > Median 0.12 0.15 0.19 0.17 0.16 0.25(0.01) (0.02) (0.02) (0.03) (0.03) (0.01)
Notes:
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification in which energy+services is flexible and raw materials is quasi-fixed. The results are estimated using a complete polynomial series of
degree 2 for each of the two nonparametric functions (G and C ) of our approach.
c. In the first three rows we report ratios of productivity for plants at various percentiles of the productivity distribution. In the remaining four rows we report estimates of the productivity differences between
plants (as a fraction) based on whether they have exported some of their output, imported intermediate inputs, spent money on advertising, and paid wages above the industry median. For example, in industry 311
for Chile a firm that advertises is, on average, 1% less productive than a firm that does not advertise.
Table D2: Heterogeneity in Productivity--Energy+Services Flexible(Structural Estimates: Gross Output)
A-21
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Labor 0.13 0.21 0.33 0.28 0.30 0.24
(0.02) (0.04) (0.03) (0.06) (0.02) (0.01)
Capital 0.05 0.07 0.02 0.01 0.05 0.06
(0.01) (0.03) (0.02) (0.04) (0.02) (0.02)
Raw Materials 0.60 0.44 0.41 0.42 0.42 0.43
(0.01) (0.01) (0.01) (0.02) (0.01) (0.01)
Energy+Services 0.22 0.30 0.23 0.26 0.28 0.29
(0.02) (0.05) (0.03) (0.07) (0.04) (0.02)
Sum 1.00 1.01 1.00 0.97 1.05 1.02
(0.01) (0.04) (0.03) (0.05) (0.04) (0.01)
Mean(Capital) /
Mean(Labor) 0.39 0.33 0.07 0.04 0.16 0.26
(0.12) (0.26) (0.06) (0.15) (0.07) (0.09)
Chile
Labor 0.19 0.34 0.35 0.38 0.42 0.26
(0.02) (0.03) (0.04) (0.04) (0.05) (0.02)
Capital 0.06 0.07 0.06 0.06 0.12 0.11
(0.01) (0.02) (0.02) (0.02) (0.03) (0.01)
Raw Materials 0.59 0.47 0.49 0.46 0.43 0.47
(0.00) (0.01) (0.01) (0.01) (0.01) (0.00)
Energy+Services 0.21 0.18 0.15 0.16 0.18 0.21
(0.02) (0.03) (0.04) (0.03) (0.06) (0.02)
Sum 1.05 1.06 1.05 1.07 1.14 1.05
(0.02) (0.02) (0.02) (0.02) (0.03) (0.01)
Mean(Capital) /
Mean(Labor) 0.32 0.20 0.16 0.16 0.28 0.41
(0.04) (0.06) (0.04) (0.06) (0.07) (0.04)
Notes:
d. The row titled "Sum" reports the sum of the average labor, capital, raw materials, and energy+services elasticities, and the row titled "Mean(Capital)/Mean(Labor)" reports the ratio of the average capital
elasticity to the average labor elasticity.
Table D3: Average Input Elasticities of Output--Raw Materials Flexible(Structural Estimates: Gross Ouput)
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification in which raw materials is flexible and energy+services is quasi-fixed. The results are estimated using a complete polynomial series of
degree 2 for each of the two nonparametric functions (G and C ) of our approach.
c. Since the input elasticities are heterogeneous across firms, we report the average input elasticities within each given industry.
A-22
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
75/25 ratio 1.24 1.27 1.25 1.24 1.24 1.31(0.02) (0.07) (0.03) (0.07) (0.07) (0.02)
90/10 ratio 1.58 1.64 1.59 1.62 1.57 1.72(0.03) (0.16) (0.08) (0.19) (0.19) (0.05)
95/5 ratio 1.92 2.10 1.90 2.01 1.89 2.13(0.06) (0.24) (0.13) (0.41) (0.30) (0.07)
Exporter 0.09 -0.01 0.04 0.05 0.01 0.04(0.04) (0.09) (0.07) (0.18) (0.11) (0.01)
Importer 0.02 0.03 0.08 -0.03 0.05 0.07(0.02) (0.09) (0.10) (0.15) (0.08) (0.01)
Advertiser -0.06 0.01 -0.01 -0.02 -0.02 -0.03(0.02) (0.05) (0.04) (0.05) (0.05) (0.01)
Wages > Median 0.04 0.11 0.14 0.09 0.11 0.13(0.02) (0.07) (0.03) (0.06) (0.07) (0.01)
Chile
75/25 ratio 1.34 1.44 1.40 1.50 1.52 1.51(0.02) (0.02) (0.02) (0.02) (0.04) (0.01)
90/10 ratio 1.82 2.13 2.05 2.31 2.26 2.32(0.07) (0.06) (0.06) (0.06) (0.13) (0.02)
95/5 ratio 2.32 2.80 2.70 3.09 2.98 3.19(0.12) (0.10) (0.10) (0.11) (0.25) (0.04)
Exporter -0.02 0.01 0.05 -0.01 -0.03 0.01(0.06) (0.03) (0.03) (0.03) (0.03) (0.01)
Importer 0.06 0.06 0.09 0.12 0.06 0.13(0.06) (0.03) (0.03) (0.04) (0.03) (0.01)
Advertiser 0.00 0.02 0.02 0.03 -0.02 0.04(0.02) (0.02) (0.03) (0.01) (0.02) (0.01)
Wages > Median 0.13 0.15 0.18 0.19 0.17 0.25(0.04) (0.03) (0.03) (0.02) (0.04) (0.01)
Notes:
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification in which raw materials is flexible and energy+services is quasi-fixed. The results are estimated using a complete polynomial series of
degree 2 for each of the two nonparametric functions (G and C ) of our approach.
c. In the first three rows we report ratios of productivity for plants at various percentiles of the productivity distribution. In the remaining four rows we report estimates of the productivity differences between
plants (as a fraction) based on whether they have exported some of their output, imported intermediate inputs, spent money on advertising, and paid wages above the industry median. For example, in industry 311
for Chile a firm that advertises is, on average, 0% less productive than a firm that does not advertise.
Table D4: Heterogeneity in Productivity--Raw Materials Flexible(Structural Estimates: Gross Output)
A-23
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Labor 0.18 0.26 0.39 0.46 0.29 0.33
(0.05) (0.07) (0.04) (0.12) (0.09) (0.02)
Capital 0.09 0.04 0.06 0.25 0.09 0.07
(0.07) (0.06) (0.04) (0.16) (0.11) (0.02)
Intermediates 0.67 0.54 0.52 0.51 0.53 0.54
(0.01) (0.01) (0.01) (0.02) (0.01) (0.00)
Sum 0.95 0.84 0.96 1.22 0.90 0.95
(0.12) (0.11) (0.07) (0.26) (0.18) (0.04)
Mean(Capital) /
Mean(Labor) 0.52 0.16 0.15 0.55 0.30 0.21
(1.42) (0.66) (0.09) (0.34) (0.35) (0.07)
Chile
Labor 0.20 0.33 0.50 0.37 0.60 0.30
(0.03) (0.07) (0.05) (0.03) (0.15) (0.02)
Capital 0.02 0.08 0.17 0.10 0.32 0.15
(0.06) (0.09) (0.07) (0.06) (0.15) (0.05)
Intermediates 0.67 0.54 0.56 0.59 0.50 0.55
(0.00) (0.01) (0.01) (0.01) (0.01) (0.00)
Sum 0.89 0.95 1.24 1.05 1.42 1.01
(0.08) (0.13) (0.11) (0.07) (0.29) (0.07)
Mean(Capital) /
Mean(Labor) 0.09 0.23 0.35 0.27 0.53 0.50
(0.25) (0.23) (0.11) (0.14) (0.35) (0.15)
Notes:
c. Since the input elasticities are heterogeneous across firms, we report the average input elasticities within each given industry.
d. The row titled "Sum" reports the sum of the average labor, capital, and intermediate input elasticities, and the row titled "Mean(Capital)/Mean(Labor)" reports the ratio of the average capital elasticity to the
average labor elasticity.
Table D5: Average Input Elasticities of Output--Fixed Effects(Structural Estimates: Gross Ouput)
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification with fixed effects and are estimated using a complete polynomial series of degree 2 for each of the two nonparametric functions (G and
C ) of our approach.
A-24
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
75/25 ratio 1.36 1.67 1.30 1.58 1.52 1.52(0.32) (0.40) (0.06) (0.46) (0.32) (0.08)
90/10 ratio 1.82 2.82 1.70 2.71 2.20 2.21(1.25) (1.71) (0.18) (2.82) (1.04) (0.25)
95/5 ratio 2.30 4.14 2.09 4.04 2.78 2.84(2.66) (4.01) (0.35) (13.90) (1.75) (0.41)
Exporter 0.26 0.25 0.10 -0.04 0.41 0.22(0.35) (0.95) (0.19) (2.64) (0.50) (0.11)
Importer 0.13 0.35 0.19 -0.08 0.32 0.27(0.29) (0.76) (0.25) (2.25) (0.37) (0.10)
Advertiser 0.01 0.32 0.07 -0.17 0.19 0.14(0.09) (0.30) (0.08) (0.41) (0.24) (0.04)
Wages > Median 0.17 0.45 0.20 0.02 0.40 0.37(0.26) (0.53) (0.06) (0.51) (0.33) (0.09)
Chile
75/25 ratio 1.57 1.60 1.52 1.52 2.06 1.57(0.15) (0.17) (0.12) (0.13) (0.36) (0.15)
90/10 ratio 2.41 2.55 2.40 2.34 4.48 2.45(0.40) (0.59) (0.45) (0.48) (1.27) (0.45)
95/5 ratio 3.14 3.38 3.38 3.20 7.30 3.41(0.61) (1.20) (0.98) (0.99) (2.55) (0.77)
Exporter 0.34 0.07 -0.07 0.07 -0.42 0.14(0.23) (0.21) (0.12) (0.42) (0.38) (0.24)
Importer 0.51 0.17 -0.02 0.22 -0.25 0.25(0.26) (0.18) (0.11) (0.31) (0.39) (0.21)
Advertiser 0.22 0.13 -0.06 0.04 -0.20 0.12(0.11) (0.14) (0.09) (0.07) (0.23) (0.10)
Wages > Median 0.50 0.28 0.10 0.23 -0.12 0.39(0.20) (0.17) (0.08) (0.15) (0.44) (0.22)
Notes:
Table D6: Heterogeneity in Productivity--Fixed Effects(Structural Estimates: Gross Output)
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification with fixed effects and are estimated using a complete polynomial series of degree 2 for each of the two nonparametric functions (G and
C ) of our approach.
c. In the first three rows we report ratios of productivity for plants at various percentiles of the productivity distribution. In the remaining four rows we report estimates of the productivity differences between
plants (as a fraction) based on whether they have exported some of their output, imported intermediate inputs, spent money on advertising, and paid wages above the industry median. For example, in industry 311
for Chile a firm that advertises is, on average, 5% more productive than a firm that does not advertise.
A-25
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Labor 0.18 0.32 0.39 0.45 0.40 0.36
(0.04) (0.04) (0.03) (0.07) (0.03) (0.01)
Capital 0.13 0.18 0.08 -0.01 0.11 0.15
(0.03) (0.02) (0.02) (0.04) (0.02) (0.01)
Intermediates 0.67 0.54 0.52 0.51 0.53 0.54
(0.01) (0.01) (0.01) (0.01) (0.01) (0.00)
Sum 0.98 1.03 0.99 0.95 1.03 1.05
(0.02) (0.03) (0.02) (0.08) (0.02) (0.01)
Mean(Capital) /
Mean(Labor) 0.72 0.55 0.21 -0.02 0.27 0.41
Chile
Labor 0.24 0.44 0.45 0.37 0.52 0.36
(0.01) (0.03) (0.02) (0.03) (0.03) (0.01)
Capital 0.12 0.11 0.07 0.09 0.14 0.17
(0.01) (0.02) (0.01) (0.02) (0.01) (0.01)
Intermediates 0.66 0.54 0.55 0.59 0.50 0.55
(0.00) (0.01) (0.01) (0.01) (0.01) (0.00)
Sum 1.02 1.09 1.08 1.04 1.16 1.08
(0.01) (0.02) (0.02) (0.02) (0.02) (0.01)
Mean(Capital) /
Mean(Labor) 0.50 0.26 0.15 0.25 0.28 0.48
(0.05) (0.05) (0.04) (0.05) (0.04) (0.02)
Notes:
c. Since the input elasticities are heterogeneous across firms, we report the average input elasticities within each given industry.
d. The row titled "Sum" reports the sum of the average labor, capital, and intermediate input elasticities, and the row titled "Mean(Capital)/Mean(Labor)" reports the ratio of the average capital elasticity to the
average labor elasticity.
Table D7: Average Input Elasticities of Output--Extra Unobservable(Structural Estimates: Gross Ouput)
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification with fixed effects and are estimated using a complete polynomial series of degree 2 for each of the two nonparametric functions (G and
C ) of our approach.
A-26
Colombia
Food Products
(311)
Textiles
(321)
Apparel
(322)
Wood Products
(331)
Fabricated Metals
(381) All
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
Gross Output
(GNR)
75/25 ratio 1.35 1.35 1.29 1.35 1.32 1.37(0.04) (0.03) (0.02) (0.08) (0.03) (0.01)
90/10 ratio 1.82 1.83 1.68 1.94 1.76 1.88(0.13) (0.08) (0.06) (0.30) (0.05) (0.02)
95/5 ratio 2.36 2.34 2.03 2.57 2.18 2.37(0.26) (0.17) (0.11) (0.70) (0.09) (0.03)
Exporter 0.15 0.03 0.04 0.32 0.11 0.06(0.07) (0.04) (0.04) (0.25) (0.04) (0.01)
Importer 0.03 0.05 0.11 0.14 0.12 0.11(0.04) (0.04) (0.04) (0.15) (0.03) (0.01)
Advertiser -0.03 0.05 0.03 0.07 0.07 0.02(0.03) (0.04) (0.03) (0.11) (0.03) (0.01)
Wages > Median 0.09 0.18 0.18 0.21 0.23 0.19(0.04) (0.04) (0.02) (0.10) (0.03) (0.01)
Chile
75/25 ratio 1.37 1.49 1.43 1.51 1.54 1.55(0.01) (0.03) (0.02) (0.02) (0.02) (0.01)
90/10 ratio 1.92 2.18 2.12 2.35 2.35 2.40(0.03) (0.09) (0.04) (0.05) (0.06) (0.02)
95/5 ratio 2.51 2.93 2.77 3.15 3.12 3.33(0.06) (0.18) (0.08) (0.11) (0.12) (0.04)
Exporter 0.00 0.03 0.09 0.00 -0.02 0.03(0.04) (0.05) (0.03) (0.04) (0.03) (0.01)
Importer 0.12 0.10 0.13 0.15 0.09 0.15(0.04) (0.04) (0.02) (0.04) (0.03) (0.01)
Advertiser 0.04 0.04 0.06 0.04 0.01 0.06(0.02) (0.03) (0.02) (0.02) (0.02) (0.01)
Wages > Median 0.20 0.19 0.22 0.22 0.20 0.30(0.03) (0.04) (0.02) (0.03) (0.03) (0.01)
Notes:
Table D8: Heterogeneity in Productivity--Extra Unobservable(Structural Estimates: Gross Output)
a. Standard errors are estimated using the bootstrap with 200 replications and are reported in parentheses below the point estimates.
b. For each industry, the numbers are based on a gross output specification with fixed effects and are estimated using a complete polynomial series of degree 2 for each of the two nonparametric functions (G and
C ) of our approach.
c. In the first three rows we report ratios of productivity for plants at various percentiles of the productivity distribution. In the remaining four rows we report estimates of the productivity differences between
plants (as a fraction) based on whether they have exported some of their output, imported intermediate inputs, spent money on advertising, and paid wages above the industry median. For example, in industry 311
for Chile a firm that advertises is, on average, 5% more productive than a firm that does not advertise.
A-27
Online Appendix O1: Monte Carlo Simulations
We consider a panel of 500 firms over 30 periods. To simplify the problem we abstract away from
labor and consider the following Cobb-Douglas production function
Yjt = Kαkjt M
αmjt e
ωjt+εjt ,
where αk = 0.25, αm = 0.65, and εjt is measurement error that is distributed N (0, 0.07). ωjt
follows an AR(1) process
ωjt = δ0 + δωjt−1 + ηjt,
where δ0 = 0.2, δ = 0.8, and ηjt ∼ N (0, 0.04). We select the variances of the errors and the AR(1)
parameters to roughly correspond to the estimates from our Chilean and Colombian datasets.
The environment facing the firms is the following. At the beginning of each period, firms
choose investment Ijt and intermediate inputsMjt. Investment determines the next period’s capital
stock via the law of motion for capital
Kjt+1 = (1− dj)Kjt + Ijt,
where dj ∈ {0.05, 0.075, 0.10, 0.125, 0.15} is the depreciation rate which is distributed uniformly
across firms. Intermediate inputs are subject to quadratic adjustment costs of the form
CMjt = 0.5b
(Mjt −Mjt−1)2
Mjt
,
where b is a parameter that indexes the level of adjustment costs, which we vary in our simulations.
Firms choose investment and intermediate inputs to maximize expected discounted profits. The
O-1
problem of the firm, written in recursive form, is thus given by
V (Kjt,Mjt−1, ωjt) = maxIjt,Mjt
PtKαkjt M
αmjt e
ωjt − P It Ijt − ρtMjt
−0.5b(Mjt −Mjt−1)2
Mjt
+ βEtV (Kjt+1,Mjt, ωjt+1)
s.t.
Kjt+1 = (1− dj)Kjt + Ijt
Ijt ≥ 0,Mjt ≥ 0
ωjt+1 = δ0 + δωjt + ηjt+1.
The price of output Pt and the price of intermediate inputs ρt are set to 1. The price of investment
P It is set to 8, and there are no other costs to investment. The discount factor is set to 0.985.
In order for our Monte Carlo simulations not to depend on the initial distributions of (k,m, ω),
we simulate each firm for a total of 200 periods, saving only the last 30 periods. The initial
conditions, k1, m0, and ω1 are drawn from the following distributions: U (11, 400) , U (11, 400) ,
and U (1, 3). Since the firm’s problem does not have an analytical solution, we solve the problem
numerically by value function iteration with an intermediate modified policy iteration with 100
steps, using a multi-linear interpolant for both the value and policy functions.49
O1.1. Inference
In the first set of Monte Carlo simulations, we provide evidence that our bootstrap procedure has
the correct coverage for our estimator. For this set of simulations, we set the adjustment cost pa-
rameter for intermediate inputs, b, to zero to correspond with our DGP. We begin by simulating
500 samples, each consisting of 500 firms over 30 periods. For each sample we nonparametrically
bootstrap the data 199 times.50 For each bootstrap replication we estimate the output elasticities of
capital and intermediate inputs using our procedure as described in Section 6. We then compute
the 95% bootstrap confidence interval using the 199 bootstrap replications. This generates 500
49See Judd (1998) for details.50See Davidson and MacKinnon (2004).
O-2
bootstrap confidence intervals, one for each sample. We then count how many times (out of 500)
the true values of the output elasticities (i.e., 0.25 and 0.65) lie within the bootstrap confidence
interval. The results are presented graphically in Figures O1.1A and O1.1B. The true value of
the elasticity is contained inside the 95% confidence interval 95.4% (capital) and 94.2% (interme-
diate inputs) of the time. Hence, for both the capital and intermediate elasticities we obtain the
correct coverage, suggesting that we can use our bootstrap procedure to do inference even in the
nonparametric case.
O1.2. Estimator Performance
For our second set of Monte Carlo simulations, we evaluate how well our estimator performs
when the first-order condition for intermediate inputs does not hold exactly. We first generate 100
Monte Carlo samples for each of 9 values of the adjustment cost parameter b, ranging from zero
adjustment costs to very large adjustment costs. For the largest value, b = 1, this would imply
that firms in our Chilean and Colombian datasets, on average, pay substantial adjustment costs
for intermediate inputs of almost 10% of the value of total gross output. For each sample we
estimate the average capital and intermediate input elasticities in two ways. As a benchmark, we
first obtain estimates using a simple version of dynamic panel with no fixed effects. The reason we
use dynamic panel is that, in light of our non-identification arguments in Section 3, this procedure
provides consistent estimates under the presence of adjustment costs. We compare these estimates
to ones obtained via our nonparametric procedure, which assumes adjustment costs of zero.
We impose the (true) Cobb-Douglas parametric form in the estimation of the dynamic panel
(but not in our nonparametric procedure) to give dynamic panel the best possible chance of recov-
ering the true parameters and to minimize the associated standard errors. Given the Cobb-Douglas
structure and the AR(1) process for productivity, we have
yjt − αkkjt − αmmjt − δ0 − δ (yjt−1 − αkkjt−1 − αmmjt−1) = ηjt − δεjt−1 + εjt.
The dynamic panel procedure estimates the parameter vector (αk, αm, δ0, δ) by forming moments
O-3
in the RHS of the equation above. Specifically we use a constant and kjt, kjt−1,mjt−1 as the
instruments.
Since the novel part of our procedure relates to the intermediate input elasticity via the first
stage, we focus on the intermediate input elasticity estimates. The comparison for the capital
elasticities is very similar. The results are presented graphically in Figures O1.2A and O1.2B. Not
surprisingly, the dynamic panel data method breaks down and becomes very unstable for small
values of adjustment costs, as these costs are insufficient to provide identifying variation via the
lags. This is reflected both in the large percentile ranges and in the fact that the average estimates
bounce around the truth. Our method on the other hand performs very well, as expected. This is the
case even though for dynamic panel we impose and exploit the constraint that the true technology
is Cobb-Douglas, whereas for our procedure we do not.
As we increase the level of adjustment costs, our nonparametric method experiences a small
upward bias relative to the truth and relative to dynamic panel, although in some cases our estimates
are quite close to those of dynamic panel. The percentile range for dynamic panel is much larger,
however. So while on average dynamic panel performs slightly better for large values of adjustment
costs, the uncertainty in the estimates is larger. Overall our procedure performs remarkably well,
even for very large values of adjustment costs. At the largest value, our average estimated elasticity
of 0.689 is less than 3 percentage points larger than the truth.
O-4
Figure O1.1A
: Monte C
arlo: Inference--Capital E
lasticityD
istribution of 95% B
ootstrap Confidence Intervals
Notes: This figure presents the results from
applying our estimator to M
onte Carlo data generated as described in O
nline Appendix O
1.1 in the absenece of adjustment costs. For each sim
ulation we
nonparametrically bootstrap the data 199 tim
es. For each bootstrap replication we estim
ate the output elasticity of capital using our procedure as described in Section 6. W
e then compute the 95%
bootstrap confidence intervals using these replications. This generates a confidence interval for each of the 500 M
onte Carlo sam
ples. In the figure we plot the low
er and upper boundaries of the confidence intervals for each M
onte Carlo sam
ple. The simulations are sorted by the m
id-point of these intervals. The true value of the elasticity is 0.25. 95.4% of the constructed confidence intervals
cover the true value.
O-5
Figure O1.1B
: Monte C
arlo: Inference--Intermediate E
lasticityD
istribution of 95% B
ootstrap Confidence Intervals
Notes: This figure presents the results from
applying our estimator to M
onte Carlo data generated as described in O
nline Appendix O
1.1 in the absenece of adjustment costs. For each sim
ulation we
nonparametrically bootstrap the data 199 tim
es. For each bootstrap replication we estim
ate the output elasticity of intermediate inputs using our procedure as described in S
ection 6. We then com
pute the 95%
bootstrap confidence intervals using these replications. This generates a confidence interval for each of the 500 Monte C
arlo samples. In the figure w
e plot the lower and upper boundaries of the
confidence intervals for each Monte C
arlo sample. The sim
ulations are sorted by the mid-point of these intervals. The true value of the elasticity is 0.65. 94.2%
of the constructed confidence intervals cover the true value.
O-6
Fig
ure O
1.2
A: M
on
te C
arlo
: Estim
ato
r P
erfo
rm
an
ce--In
term
ed
iate
Inp
ut E
lastic
ityG
NR
an
d D
yn
am
ic Pan
el: Av
erages
No
tes: This fig
ure p
resents th
e results fro
m ap
plyin
g b
oth
ou
r estimato
r and
a dyn
amic p
anel d
ata estimato
r to M
onte C
arlo d
ata gen
erated as d
escribed
in O
nlin
e Ap
pen
dix
O1
.2. T
he d
ata are gen
erated su
ch th
at the first o
rder co
nd
ition n
o lo
nger
ho
lds b
ecause o
f qu
adratic ad
justm
ent co
sts. The p
arameter b
ind
exes th
e deg
ree of ad
justm
ent co
sts in in
termed
iate inp
uts facin
g th
e firm, w
ith h
igher v
alues rep
resentin
g larg
er adju
stmen
t costs. T
he y-ax
is measu
res the av
erage estim
ated
interm
ediate in
pu
t elasticity for b
oth
estimato
rs across 1
00
Mo
nte C
arlo sim
ulatio
ns. T
he tru
e valu
e of th
e averag
e elasticity is 0.6
5.
0.6
0.6
5
0.7
0.7
5
0.8
0.8
5
0.0
00
.10
0.2
00
.30
0.4
00
.50
0.6
00
.70
0.8
00
.90
1.0
0
Average Intermediate Input Elasticity
Valu
e of A
dju
stme
nt C
ost P
arameter (b
)
Dyn
amic P
anel
GN
R
O-7
Fig
ure O
1.2
B: M
on
te Ca
rlo: E
stima
tor P
erform
an
ce--Interm
edia
te Inp
ut E
lasticity
GN
R a
nd
Dy
na
mic P
an
el: Per
cen
tile Ra
ng
es
No
tes: This fig
ure p
resents th
e results fro
m ap
plyin
g b
oth
ou
r estimato
r and
a dyn
amic p
anel d
ata estimato
r to M
onte C
arlo d
ata gen
erated as d
escribed
in O
nlin
e Ap
pen
dix
O1
.2. T
he d
ata are gen
erated su
ch th
at the first o
rder co
nd
ition n
o lo
nger h
old
s
becau
se of q
uad
ratic adju
stmen
t costs. T
he p
arameter b
ind
exes th
e deg
ree of ad
justm
ent co
sts in in
termed
iate inp
uts facin
g th
e firm, w
ith h
igher v
alues rep
resentin
g larg
er adju
stmen
t costs. T
he y-ax
is measu
res the 2
.5 an
d 9
7.5
percen
tiles of th
e estimated
interm
ediate in
pu
t elasticity for b
oth
estimato
rs across 1
00
Mo
nte C
arlo sim
ulatio
ns. T
he tru
e valu
e of th
e averag
e elasticity is 0.6
5.
-0.6
-0.4
-0.2 0
0.2
0.4
0.6
0.8 1
1.2
0.0
00
.10
0.2
00
.30
0.4
00
.50
0.6
00
.70
0.8
00
.90
1.0
0
Average Intermediate Input Elasticity
Valu
e o
f Ad
justm
en
t Co
st Param
eter (b
)
Dyn
amic P
anel (9
7.5
percen
tile)
GN
R (9
7.5
pe
rcen
tile)
Dyn
amic P
anel (2
.5 p
ercentile
)G
NR
(2.5
pe
rcen
tile)
O-8
Online Appendix O2: Value Added
In this appendix we provide additional details regarding value added.
Restricted Profit Functions
Recall equation (24) in the main body:
V AEt = F (kt, lt,mt) eωt −Mt≡ V (kt, lt, e
ωt) .
It can be shown that the total derivative of value added with respect to one of its inputs is equal to
the partial derivative of gross output with respect to that input. For example, the total derivative of
value added with respect to productivity is given by:
dV AEtdeωt
=dV (kt, lt, e
ωt)
deωt
=
[∂F (kt, lt,mt) e
ωt
∂eωt−(∂F (kt, lt,mt) e
ωt
∂Mt
− 1
)∂Mt
∂eωt
]=
∂GOt
∂eωt.
Due to the first order condition in equation (14) in the main text, the term inside the parentheses
on the second line is equal to zero, where the relative price of output to intermediate inputs has
been normalized to one via deflation. This implies that:
(∂GOt
∂eωteωt
GOt
)︸ ︷︷ ︸
elasGOteωt
=
(dV AEtdeωt
eωt
V AEt
)︸ ︷︷ ︸
elasV Ateωt
V AEtGOE
t
=
(dV AEtdeωt
eωt
V AEt
)︸ ︷︷ ︸
elasV Ateωt
(1− St) .
However, once we add back in the ex-post shocks we have the following:
dV AEtdeωt
=dV (kt, lt, e
ωt , eεt)
deωt
=
[∂F (kt, lt,mt) e
ωt+εt
∂eωt−(∂F (kt, lt,mt) e
ωt+εt
∂Mt
− 1
)∂Mt
∂eωt
].
O-9
Notice now that the term inside the parentheses is no longer equal to zero, due to the presence of
the ex-post shock, εt. The reason is that the first order condition, which previously made that term
equal to zero, is an ex-ante object, whereas what is inside the parentheses is ex-post. Therefore,
we cannot simply transform the value added elasticities into their gross output counterparts by
re-scaling via the ratio of value added to gross output.
The first order condition implies that ∂F (kt,lt,mt)eωt+εt
∂Mt= eεt
E . In turn, this implies that
dV AEtdeωt
=
[∂F (kt, lt,mt) e
ωt+εt
∂eωt−(eεt
E− 1
)∂Mt
∂eωt
]⇒ elas
V AEteωt = elasGOteωt
GOt
V AEt− ∂Mt
∂eωt+εteωt
V AEt
(eεt
E− 1
).
The equation above can then be rearranged to form relationship between the elasticities as:
(∂GOt
∂eωteωttGOt
)︸ ︷︷ ︸
elasGOteωt
=
(∂V AEt∂eωt
eωt
V AEt
)︸ ︷︷ ︸
elasV AEteωt
(1− St) +∂Mt
∂eωteωt
GOt
(eεt
E− 1
).
A similar result holds when we analyze the elasticities with respect to the entire productivity shock,
eωt+εt , instead of just the persistent component, eωt . In this case we have the following relationship:
(∂GOt
∂eωt+εteωt+εt
GOt
)︸ ︷︷ ︸
elasGOt
eωt+εt
=
(∂V AEt∂eωt+εt
eωt+εt
V AEt
)︸ ︷︷ ︸
elasV AEt
eωt+εt
(1− St) +∂Mt
∂eωt+εteωt+εt
GOt
(eεt
E− 1
).
“Structural” Value Added
As discussed in Section 8.2, for the Leontief case we have
Yt = min [H(kt, lt), C (mt)] eωt+εt . (56)
The standard Leontief condition,H(kt, lt) = C (mt), will not generally hold unless C (mt) = aMt.
Even in this linear case, the value added production function, H(kt, lt)eωt+εt , does not relate
cleanly to the empirical measure of value added V AEt ≡ Yt−Mt, since V AEt = H(kt, lt)(eωt+εt − 1
a
).
O-10
However, it does correspond directly to gross output since
Yt = H(kt, lt)eωt+εt .
Neither of these issues are resolved by moving ωt inside the min function in equation (56). Sup-
pose that instead of equation (56), one wrote the production function as: Yt = min[H (kt, lt) eωt ,
C (mt)]eεt . For similar reasons, the condition, H(kt, lt)e
ωt = C (mt) , only holds when C (mt) =
aMt. Even when this is the case, the value added production function will again not correspond
to the empirical measure of value added as V AEt = H(kt, lt)eωt(eεt − 1
a
). As was the case above,
however, it directly corresponds to gross output: Yt = H(kt, lt)eωt+εt .
It is also the case that moving εt inside the min function does not help. The problem is that the
key condition,H(kt, lt)eεt = aMt, will not hold when ωt is outside the min because of the presence
of the ex-post shock εt. Since εt is realized after input decisions are made, the key condition will
generally not hold. Thus neither V AEt , nor gross output Yt correspond to the structural value-
added production function H (kt, lt) eωt+εt . An analogous argument holds when ωt is inside the
min function.
References Online AppendixDavidson, Russell and James G MacKinnon. 2004. Econometric Theory and Methods, vol. 5. New
York: Oxford University Press.
Judd, Kenneth L. 1998. Numerical Methods in Economics. Cambridge: MIT Press.
O-11