HOHENHEIMER

DISKUSSIONSBEITRÄGE

Are Less Constrained Governments Really More Successful in Executing

Market-oriented Policy Changes

von

Hans Pitlik

Nr. 255/2005

Institut für Volkswirtschaftslehre (520)

Universität Hohenheim, 70593 Stuttgart

ISSN 0930-8334

Are Less Constrained Governments Really More

Successful in Executing Market-oriented Policy

Changes?

Hans Pitlik* University of Hohenheim, Department of Economics 520D

70593 Stuttgart, Germany pitlikha@uni-hohenheim.de

Abstract

It is commonly suspected that market-oriented reforms require a centralized and autonomous

government. According to this view, institutional constraints on a government weaken

decisiveness of state action and constitute a major obstacle to policy reform. However, this

line of reasoning neglects the importance of government credibility for implementation and

consolidation of policy reforms. Lack of credibility causes higher economic and political costs

of reform and reduces incentives of governments to undertake risky policy changes. Contrary

to conventional wisdom, more veto gates may at the same time increase both government

credibility and decisiveness, and may therefore be conducive to reform. This alternative view

of the role of institutional constraints is supported by empirical evidence.

JEL classification: D78, P11, P21

Keywords: democracy, market-oriented reform, political credibility, veto players

* The author is especially indebted to Friedrich Heinemann, Sebastian Moll, Jörg Naeve, and the participants and discussants at DISS.KURS at the University of Hohenheim and at the ZEW Seminar, Mannheim.

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1 Introduction

The questions of why, when, and under which circumstances, governments engage in market-

oriented policy reforms has attracted enormous interest in the political and economic literature

during the past twenty years. The Political Economy of Reform, surveyed by Rodrik (1996),

Sturzenegger and Tommasi (1998), and Drazen (2000), has produced two central ideas,

generally based on game-theoretic modeling. On the one hand, it is commonly agreed upon

that economic crises make market-friendly reforms of economic policies more likely. Crises

make past policy failures visible, may help breaking interest group resistance to reform, and

may create a sense of urgency that appears to be conducive to a large-scale policy change. A

number of recent investigations (Lora, 1998, Drazen and Easterly, 2001, Pitlik and Wirth,

2003, and Heinemann, 2004) find empirical evidence in favor of the crisis hypothesis.

On the other hand, it is often claimed that a more autonomous government (not to say: a

non-democratic regime) that does not have to deal with political veto players (not to say:

voters), is good for reform. A government less constrained by checks and balances appears to

be more decisive as it does not depend on the consent of numerous reform-opposing groups

and their representatives (Cox and McCubbins, 2001, Tsebelis, 2002). Reform-minded

autonomous governments may even be able to completely disregard special interest groups'

and a general public's antagonism to market-oriented policy changes. Yet, as regards far-

reaching economic policy reforms support of the autonomy hypothesis is at best mixed. In his

studies, Tsebelis (2002) finds that increasing the number of veto players in a polity reduces

the decisiveness of political action in diverse policy fields. Veiga (2000) observes that less

democratic and less fragmented governments tend to stabilize earlier from high inflation

episodes. Both Veiga's and Tsebelis' analyses are however restricted to rather limited country

samples. Contrary to the autonomy hypothesis, Fidrmuc (2003) reports that democracy has

proven favorably for economic reform in post-communist transition economies. De Haan and

Sturm (2003) validate this result for a set of developing countries. Pitlik and Wirth (2003)

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also reveal a positive relation between overall economic liberalization and constrained

government in a broad sample of both developing and industrial countries.

However, empirical contributions which challenge the autonomy hypothesis usually do

not offer a concise explanation for their findings. A standard answer is that democracy may be

a superior mechanism of conflict resolution. If this is correct, it would hardly be surprising

that democratic governments are more successful in executing market-oriented policy changes

as these are almost always characterized by severe political conflicts between prospective

losers and prospective winners of economic reforms. But how do these assertions relate to

standard theoretical thinking in the Political Economy of Reform? As shall be argued in the

next two sections of this paper, the missing link might be found in a re-interpretation of

reform theory in the light of credibility problems during the implementation of policy

changes.

Another open question is how much we can be confident in empirical results. Findings

in Political Economy are often driven by methodology. Recent availability of economic and

political cross-section/time-series data for an increasing number of countries and an extended

period of time have made the use of panel data methods almost a standard tool. In order to

deal with unobserved heterogeneity which is nearly always a problem when it comes to

explaining differences in economic and political variables, modern econometrics recommends

researchers to include unit fixed effects. Yet, employing unit fixed effects also has its cost in

terms of lost cross-sectional information, especially when (almost) time invariant political

characteristics of a unit serve as explanatory variables. Loss of inference on (almost) time

invariant political variables may outweigh gains of modeling heterogeneity. No method

qualifies as the single best way to estimate the effects of political structures on observed

behavior. To increase confidence in results one might be best advised to compare results from

different estimation techniques as a second best-solution. Papers mentioned above, however,

usually refrain from this. Hence, this paper also reports results of new estimates based on an

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extended data set and several regression techniques. As shall be shown, there is ample

evidence in favor of the hypothesis that constrained governments are more successful in

executing market-oriented policy changes.

2 A brief look at standard theory, and a re-interpretation

Standard explanations of market-oriented policy changes generally start with the puzzle: 'If

reforms are so beneficial to society as a whole, why do we observe so little reform activity?'

Then, in a next step several political impediments to beneficial reforms are recognized, and in

a third step the conditions under which these barriers can be evaded are identified. This

section briefly reviews the most acknowledged theories of reform obstacles and aims at a new

interpretation in terms of government credibility.

2.1 Political myopia

Reform aversion is often explained by political myopia. In short, the story goes as follows:

Governments dislike market-friendly policy changes since the economic benefits of reforms

show up only in the longer run whereas adjustment costs must be borne today. Politicians

motivated by re-election have no incentive to initiate a policy change as they fear being voted

out of office due to a temporary economic downturn (Garrett and Lange, 1996). The higher

the probability of losing the next election, the more myopic politicians tend to act (Alesina

and Tabellini, 1990). Autocracies therefore appear to be favorable for reforms as dictators do

not have to direct attention on the economic situation on election day.

Yet, this overlooks that autocratic rulers can be thrown out of office, too (Tullock,

1987). It is by no means clear that dictators have a longer time horizon than democratic rulers

as they can suddenly lose power by an unpredicted coup d'état (Clague et al., 2001).

Moreover, democratic politicians' shortsightedness is only reasonable if voters punish them

even for a temporary deterioration. Thus, it is voters' myopia which may be the main obstacle

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to reform. But why should rational voters oppose a reform that benefits a majority of them in

the long run? A possible explanation may be lack of knowledge. However, it seems to be

unsound to argue that citizens as market actors behave in their long run economic interests,

while in their role as voters they act myopically (Wittman, 1995). To be sure, there is some

evidence for "rational irrationality" (Caplan, 2001, Heinemann, 2001) and ideological status

quo bias in politics. Crises may then be conducive to a policy change as they induce learning

about the correct model of the world.

Brennan and Buchanan (1985) offer an alternative explanation for rational voter

myopia. In their view, an inherent uncertainty in politics makes voters more likely to neglect

the longer run consequences of current democratic decisions. As long as voters cannot be sure

to enjoy the future benefits of reforms which result from today's sacrifices they prefer smaller

short run benefits of a current policy even if it is unsustainable in the long run. Changing

political majorities mixed with unlimited government are thus at the core of this line of

reasoning. Myopic voting behavior is a consequence of a lack of credible political

commitment not to exploit the future benefits of a policy change.

This logic parallels theories of private investment in uncertain future environments.

Rodrik (1991) shows that investment depends critically on expectations about future policies.

If investors believe that a market-oriented policy change will be (partly) reversed it is often

sound to wait in order to avoid sunk cost. However, large and rapid new investment is often

crucial for economic success of a policy reform. Risks of a reform reversal may generate

insufficient investment which in turn reduces the probability of a successful reform, thus

causing a self fulfilling-prophecy of reform failure. Hence, economic adjustment cost to a

reform largely depend on the credibility of the government, a fact which is well known from

monetary policy (Barro and Gordon, 1983). A less credible government that worries more

about current reform cost may therefore be less willing to initiate a policy change.

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2.2 Individual uncertainty about reform benefits

In a frequently cited paper Fernandez and Rodrik (1991) stress the role of individual-specific

uncertainty in the non-adoption of beneficial reforms. Policy changes almost always create

both winners and losers. Yet, while for some groups it is obvious that their members will

benefit from a policy change, and for others it is clear that members will lose, it is often the

case that ex ante identical members of a particular group are individually uncertain about the

effects of a specific reform on their well-being. Risk-neutral individuals base their decision to

support or to oppose reform on the expected benefits of a policy change. Fernandez and

Rodrik (1991) demonstrate that when no single group forms a political majority by itself, a

reform which would eventually benefit a majority of the population may be rejected by a

majority ex ante, although voters are aware that ex post a majority is gaining from its

adoption. The crucial point is that under these circumstances the pivotal voter comes from the

group with uncertain pay offs. If expected benefits of this group's members are negative, they

form a coalition with prospective losers and block reform.1 According to this logic even a

temporary insulation of a reform-minded government from voter pressure might be sufficient

for reform. Reforms will not be reversed if a democratic government is set in place, then,

since after reform adoption a majority of the population would vote against reversal.

An alternative route to reform in this setting would be for prospective winners to offer

compensations to potential losers. If total gains from a policy change exceed total losses, as is

assumed, winners may well be able to compensate losers in order to buy reform support. Yet,

compensation contracts are threatened by a serious inter-temporal commitment problem

(Dixit and Londregan, 1995, Acemoglu and Robinson, 2001). After reform implementation

winners from a reform built a political majority that may simply renege on a compensation

contract. Hence, ex ante no potential loser is willing to accept a compensation promise for the

future. Paying off prospective losers prior to adoption of reform, however, just turns the

problem on its head. The group of sure winners will not offer compensation because the

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beneficiaries cannot credibly commit to support policy change later. Moreover, as a fraction

of the decisive group are not identified ex ante as winners, supply of compensation may fall

short of demand. Compensation schemes fail if partners in a political deal cannot credibly

commit to keep promises. Note that this failure does not depend on asymmetric information

about reform gains and losses which may further impede implementation of compensation

schemes (Dewatripont and Roland, 1992).

2.3 Reform as a collective good

Since Olson's seminal contributions (Olson, 1965, 1982) it is widely acknowledged that

collective action problems are at the heart of political economy considerations. The success of

interest groups in obtaining special favors rests largely on the inability of the general public to

organize political pressure in favor of reform. Accepting this, the political decision making

process is often recognized as a game whose main actors are interest groups and their

representatives in the legislative, the bureaucracy, and in the executive. Seen from this view,

the utmost difficulty of market-oriented reform implementation is to induce pressure groups

to give up their special privileges. Groups however resist the elimination of tax preferences,

subsidies, regulations, import tariffs, etc., as the cost must be borne entirely by their members

while economic benefits disperse to numerous other actors. On the whole it would yet be

beneficial for all to completely ban all privileges. Put differently, policy reform becomes a

vast collective action game among interest groups, each one preferring to free ride on reform

efforts by other groups. The dilemma is more severe, the more groups participate in the

political game. In the case of fiscal policy this causes the famous common pool problem, see,

e.g., Weingast et al. (1981). Attractiveness of autocratic rule, or unrestricted majority rule,

stems from the implicitly assumed power of governments to force 'uncooperative' groups into

a 'cooperative' agreement (Haggard, 2000, p. 37). Repeated interaction among groups may

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surely generate a voluntary agreement to a mutual beneficial reform, too. Cooperation,

however, depends on the availability of a credible enforcement technology.

The role of crises in a dynamic collective action game has been analyzed in the context

of fiscal policy by Velasco (1998). Velasco assumes that a crisis leads to a deterioration of the

groups' pay offs in a non-cooperative, i.e. non-reform, equilibrium. At some critical point the

nature of the game switches from a prisoners' dilemma to an assurance game in which

cooperation (i.e. giving up privileges) is preferred by each group to non-cooperation, provided

that all other groups behave cooperatively too. In Velasco's full information, two player game

version reform comes about immediately. Yet, if the game is extended to more than two

players and if one allows for private information, switching to a superior reform equilibrium

requires a critical mass of cooperating players (Pitlik, 2003). No group is willing to give in

unless members are confident that at least a critical number of other groups also concede. A

policy change will thus be arrived at with higher probability if political actors can credibly

commit to cooperative behavior.

2.4 Veto players and a war of attrition

A central theme in the Political Economy of Reform is the notion that as the effective number

of political veto players increases, a polity becomes less decisive. Providing more actors with

diverse interests with a right to block policy change increases political transaction cost and

reduces the potential for reform (Buchanan and Tullock, 1962, Cox and McCubbins, 2001,

Tsebelis, 2002). The number of political veto actors increases, e.g., with the number of

partisan interests in a coalition government, diverse party majorities in a bicameral legislature,

or a constitutional separation of powers. A policy reform can be legally passed only if all

political players agree. Reaching consensus is however more complicated, the higher the

number of actors who have to concur. From this assertion it appears to follow directly that an

unconstrained executive that does not have to rely on the consent of other political actors is

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superior when it comes to enacting market-friendly policy reforms. In particular in the case of

an economic crisis, when urgent political action is required, a lower number of veto players in

a polity seems to be advantageous.

In line with this reasoning, Alesina and Drazen (1991) analyze interactions of two veto

actors (interest groups) in a dynamic policy reform game. In this 'war of attrition', groups

engage in a bargaining procedure over the distribution of reform costs. It is assumed that the

player who concedes first in the bargaining game bears a higher fraction of total reform costs.

This generates an incentive for each player not to concede too soon, i.e. an incentive to block

reform, hoping that the other player will give in first. As a consequence, mutual beneficial

policy changes are enacted only after a longer period of political inaction. Reform occurs

when the weakest player – whose identity is not commonly known ex ante – 'throws the

towel'. If, on the contrary, a country's political institutions do not provide several groups the

right to veto policy changes, reform takes place earlier. Also, a severe crisis reduces the non-

reform pay offs in the game significantly and therefore makes a reform delay less likely.

The crucial point here is that the timing of a reform depends on a predetermined fraction

of reform cost, the loser in this game has to accept. The more equal the cost sharing scheme

is, the earlier an agreement will be achieved. However, Alesina and Drazen (1991) are not

quite clear about the political mechanisms that may determine an unequal distribution of

reform cost. The authors claim that a disproportional sharing of reform cost may be owed to a

lack of political cohesion. This is hardly plausible, yet. In principle, there is no reason to

suspect that lack of cohesion causes an unequal division of total reform cost. Even the most

heterogeneous actors may agree on a fair cost sharing scheme ex ante. Disproportional

division of the burden may however be a consequence of an unlimited opportunity of the

'winners' to shift a major fraction of reform cost to the group of 'losers'. If the risk of ex post-

exploitation is low, there is not much to gain from blocking reform. Overcoming obstacles to

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a beneficial policy change again crucially depends on the ability of actors to credibly commit

not to break agreements.

3 The relation between political regime type, decisiveness and credibility

So far it has been argued that credibility plays a crucial role both for economic and political

reform cost. In the economic sphere, lack of credible commitment not to reverse reforms

increases adjustment cost and makes reform success less likely. In the political sphere, lack of

credible commitment of the incumbent government to respect arrangements to compensate

losers, or not to exploit minorities, raises the political cost of reform. To be more precise: if a

ruler is not believed to keep political promises, resistance to policy change is stronger because

prospective losers have more to lose. Distributional conflict, which is the source of opposition

to beneficial reform, becomes less severe if reform-minded governments can credibly commit

to adhere to political contracts. Keefer (2004, p. 1), additionally notes that a higher credibility

may also encourage governments to carry out reforms. In particular, he argues that rulers

knowing that "their promises regarding the future are not credible have less incentive to

undertake policies that only bear fruit if citizens believe government promises regarding the

future." As a lack of credibility reduces benefits of reform and increases the cost of reform,

these governments shy away from a policy change.

But which factors determine government credibility? An exhaustive game-theoretic

literature shows that credibility is closely related to reputation, i.e., "the idea … that if the

player always plays in the same way, his opponents will come to expect him to play that way

in the future" (Fudenberg and Tirole, 1991, p. 367). North and Weingast (1989), as well as

Clague et al. (1996), argue that it is nearly impossible to build up reputational capital in

autocracies, as any political promise is exclusively linked to the political fate of the ruler. A

number of recent studies by Dixit et al. (2000), Rodrik (2000), de Figuereido (2002), and

Dixit (2003) show in an infinitely repeated game framework that it may be perfectly rational

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even for unconstrained political rulers to refuse to exploit political rivals and to stick to a

moderate policy of political compromise if the following three conditions hold:

(1) Political actors do not discount future benefits of being in office and costs of being out

of office too heavily;

(2) there is a high enough probability that a ruler in power will not stay in power forever;

(3) there is a high enough probability that a political actor, once out of power, will return to

power not too far in the future.

While conditions (1) and (2) may be questionable for autocratic rulers, condition (3) is clearly

almost always not fulfilled in autocratic regimes. In democracies, the institutions of political

parties with a longer time horizon than individual politicians, and regularly held elections

contribute to all three requirements. Thus, in general we would expect a higher credibility of

democratic policies as compared to autocratic rule, per se.

All democracies, however, are not all alike. It has been stated above that several types

of democratic regimes differ by the number of formal veto actors. It is debatable whether

presidential or parliamentary regimes, unicameral or bicameral systems, federal or unitary

states, or countries with majority rule or proportional representation, are a priori characterized

by more or less effective veto players (Cox and McCubbins, 2001, Tsebelis, 2002). Yet, an

increasing number of effective veto players, who have the right to block both a desirable and

an unwanted policy change, unambiguously contributes to political resoluteness, i.e. "the

ability … to commit to maintaining a given policy" (Cox and McCubbins, 2001, pp. 26-27).

From a static perspective, higher transaction costs of political change enhance credibility, but

decrease decisiveness as well. There seems to be a deep-seated trade off between decisiveness

and resoluteness of a polity, depending on the number of veto gates (Levy and Spiller, 1994,

Cox and McCubbins, 2001). From a dynamic perspective, as shall be argued soon below, this

is far from being clear.

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Before returning to that question, a few words should be said about a third way of

increasing credibility of a policy often mentioned in the literature, i.e. delegation. Some

scholars argue that a delegation of competencies to politically independent agencies, e.g. an

independent central bank, is a promising means for governments to credibly commit to a

policy (Rogoff, 1985). Insulation from of voter and interest groups influence enables

independent agencies to enact policies that are beneficial for the economy as a whole in the

long run, but run against day-to-day political motives. Dixit (1996) and Posen (1998) however

argue that delegation does not solve the problem of missing credibility per se, as an agency's

decision could always be overrun by the delegating political actors. If a government cannot

credibly commit to a certain policy, it can also not commit to respect a new agency's

'independence', too. But if a government is able to commit not to override a new agency's

policy, it can also commit to the policy. Confirming this view, Moser (1999), and Keefer and

Stasavage (2003) observe that central bank independence has its positive effect on price

stability only if the number of political veto players is high.

The notion of political credibility is related to the irreversibility of policy decisions. But

does that necessarily mean that a multiplicity of veto players always comes at the cost of

lower decisiveness, as contributions cited above do suggest? In a static perspective, a

government unconstrained by veto actors may of course be able to execute economic policies

which face strong opposition of interest groups harmed by that policy. Yet, there are strong

reasons to presume that the political cost of reform are much higher in a dynamic perspective

if a new policy is based on command rather than compromise. In this respect, Stiglitz (1998,

pp. 19-20) notes that

"in a consensus-based system, an issue is over when everyone has come to a mutually

acceptable agreement. Because the process by which a decision is made is viewed to be

fair, even the 'losers' feel committed to upholding it; and because the consensus process

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typically provides some accommodation to all parties, there is a sense in which no one

need feel defined as a loser. As a result, once an issue has been decided, it is likely that

the issue will stay closed, at least until a major change in the world occurs. In an

adversarial system issues are never closed."

Negotiation over reform, which is a necessary condition in political systems with more than

one veto player, allows a bundling of policies in different areas. In a simple model, including

a government and two veto groups, Martinelli and Tommasi (1997) show that bundling may

be the only feasible way for government to break reform deadlock. If a reform package as a

whole has a positive net outcome for political actors, but separately enacted policy changes

would be vetoed as each would harm one of the groups, a kind of implicit log rolling-scheme

is needed for reform. To be sure, log rolling is feasible in an unconstrained majority system as

well. Yet, cycling and cheating are major problems for the stability of log rolling-coalitions,

especially in an inter-temporal context (Bernholz, 1973, Weingast and Marshall, 1988). If the

players have a right to veto a future policy change, the risk of being exploited by a new log

rolling-coalition is significantly reduced. But if cycling is reduced we should not observe a

never-ending process of forming new coalitions which is detrimental to decisiveness.

In the same vein, Spiller et al. (2003, p. 24) conjecture that an independent judiciary

increases decisiveness. Knowing that a future violation of a contract will be penalized by the

judiciary, e.g. political compensation schemes are more credible. Opposition to reform is

mitigated, thus providing a government more room for political action and thereby increasing

decisiveness of a polity.

To sum up, interrelations between decisiveness and political credibility (resoluteness)

may not be characterized by a simple trade off. While an increasing number of veto actors in a

polity increases the credibility of a reform policy, it may also improve decisiveness by

alleviating political conflict in the reform process.

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4 Empirical evidence

Summing up theoretical reasoning leads to the following two complementary hypotheses with

respect to the influence of political institutions on market-oriented reform activities.

Hypothesis 1:

Democracies are more successful in executing reforms than their autocratic counterparts.

Hypothesis 2:

Countries with higher political veto gates for an executive are more successful in executing

reforms.

4.1 Data

A first empirical difficulty now is to find suitable measures of economic reform policies and

of the political-institutional framework. The dependent variable in the present investigation is

the change in the aggregate Economic Freedom of the World-index (efw), made available by

the Fraser-Institute (Gwartney and Lawson (2002)). It covers a time span of 30 years from

1970 to 2000 in intervals of five years. The efw-index quantifies the degree of economic

freedom in a country on a scale from 0 (unfree) to 10 (totally free). It is calculated by

employing, for the most part, objective measures of market-friendliness of a nation's

economic policy. Starting from a total of 37 sub-categories, variables are grouped into five

important policy areas (size of government, security of property rights, sound money and

price stability, freedom to exchange with foreigners, and regulation of credit, labor and

business). The overall index is an unweighted average of its components. Hence, an increase

of the efw-score over a five year period (lib = ∆ efw) signals a general increase economic

freedom, i.e. a liberalization of economic policies on the whole. In its most recent edition the

efw-index covers a sample of 123 countries from regions all over the world, starting with 59

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observations in 1970. For the purpose of this study, a chain-weighted index, which corrects

for limited data availability for some of the sub-components over time, is employed.

According to the efw-index, in 2000, the most economically free nations are Hong Kong

(8.8) and Singapore (8.6), followed by the U.S. (8.5), the U.K. (8.4), Switzerland and New

Zealand (both 8.2). The most unfree nations in 2000 are the Democratic Republic of Congo

(3.7) and Myanmar (3.4). In the entire sample, the most significant reform events during a

five year interval as measured by a change (lib) in the efw-index are observed in Peru (1990-

95, +2.7), El Salvador (1990-95, +2.6), and the Czech Republic (1990-95, +2.5). Over a 20

year period from 1980 to 2000, the highest increase of the efw-index is recognized for Uganda

(+3.9). These statistics are, however, not too surprising if one considers that these countries

initially faced significant restrictions of economic freedom in the period preceding reforms.

Put differently, countries that have already realized a high degree of economic liberty do not

have the potential (nor the incentive) to further liberalize economic policies. Hence, in

regressions it will be controlled for the starting level of the efw-index.

Among the explanatory variables which represent institutional restrictions on the

executive, a nation's degree of democracy is measured by polity, which is an aggregate

indicator of political liberties from the Polity IV-project (Marshall and Jaggers, 2000). The

polity-index describes the existence and fairness of the voting process as observed by country

experts. For the purpose of this paper, the polity-index is re-coded on a 0-1-scale, with higher

values representing a higher degree of democracy. As data are available on an annual basis,

averages of polity over the respective time period will be used. Most western democracies

arrive at a polity-score of 1 throughout the period from1970 to 2000.

Restrictions on executive behavior are indicated by polconv, calculated by Henisz

(2000). This variable measures the degree of institutional constraints on the executive. It is

derived from a simple spatial model of politics including five possible veto points: the

executive, up to two legislative chambers, an independent judiciary and the existence of

16

autonomous sub-central governments. In calculating the polconv-index, Henisz also takes into

account the fragmentation of legislatures and diverting policy preferences of veto actors. The

index potentially ranges from 0 to 1, higher values indicating more severe restrictions on the

executive to change policies autonomously. In contrast to the democracy-index polity,

polconv shows a little more variablity over time, as its value depends on specific political

conditions in a country, e.g., whether it is ruled by a legislative majority in both chambers or

by a divided legislature. Yet, in established democracies polconv is still very stable.

As an alternative to polconv, the variable checks is employed, taken from the Database

of Political Institutions (Beck et al., 2001). This variable calculates the effective number of

veto players by counting the number of autonomous political parties controlling the executive

and the legislature. In contrast to polconv, checks does not count the judiciary or federalism as

potential veto points. In autocracies, the number of veto actors is assumed to be one. In

democratic regimes, the number of effective veto players is adjusted if there are important

ideological differences between governing parties, by simply adding fictional players. As

marginal effects of more and more veto actors on a government's ability to act independently

decline, checks is used in a logarithmic specification. Both polconv and checks are measured

by the mean over the respective period. All variables measuring institutional restrictions are

expected to show positive coefficients in the regressions.

Government credibility depends to a large extent on the trust of respective actors that

future promises are kept. If the executive is replaced often times, this may diminish both

incentives to follow a long run economic policy conception as well as confidence in

resoluteness of a polity. To capture these effects, govchg measures the average number of

changes of the head of the executive. In democracies, govchg only counts changes when a

change of the head of the executive is also accompanied by a change of the ruling party's

ideology. In autocratic regimes, every change of a political ruler is counted. The expected

17

effect of govchg on reform is negative. Data are derived from the Database of Political

Institutions (Beck et al., 2001).

To control for effects of crises on economic liberalization the (log of) average inflation

rates (inflation) and average GDP growth rates (growth) are included. Variables are taken

from World Bank (2001) statistics and will be lagged one period. Whereas inflation is

expected to be positively related to economic reform, estimated coefficients of growth should

be negatively related to reform. Finally, it is controlled for trade openness of the economy and

for the state of education of the population. Trade openness (open) is simply the relation of

the sum of exports plus imports to GDP. Education is proxied by the (log) ratio of secondary

school enrolment (secenrol). Both variables are obtained from World Bank (2001). As more

open economies might face a larger outside pressure to liberalize (Sinn, 1997), and as better

educated people might be more competent in judging inadequate policies (Heinemann, 2004)

a positive sign of the respective coefficients is expected. Summary statistics and a correlation

matrix of all variables employed in the analysis are reported in the appendix.

4.2 Methods

A second empirical problem is to find suitable techniques in order to estimate the effects of

democracy and veto players on economic liberalization. The general model to be estimated is

given by

t,i1t,i1t,it,it,i1t,it,i uXscrgovchgexrestrefwlib +′⋅µ+′⋅δ+⋅γ+⋅β+⋅α= −−− , (1)

where i represents country units, t represents a respective time period, exrestr is one of the

institutional variables (polity, polconv, checks) for contraints on the executive, govchg

18

represents the number of government changes, crs' is a vector of crisis variables (inflation,

growth), and X' is a vector of further controls (open, secenrol).

The term u can be decomposed to u , with unit fixed effects ,

time fixed effects µ , and an error term . There are of course good reasons to include both

time and unit fixed effects in an analysis of political and economic determinants of reform.

Time fixed effects control for (unknown) external reform pressures which may affect all

countries similarly. One may think, e.g., of a worldwide spread of economic ideas leading

governments in any part of the world to liberalize their economic policies. Inclusion of unit

fixed effects can be justified by unobserved country heterogeneity. There is such a variety of

conceivable country-specific influences on politics that it is impossible to formulate a fully

specified model. Country dummies may help capturing a systematic influence of omitted

variables, at least to a certain degree.

t,i

t

t,itit,i ε+µ+η= iη

t,iε

Still the most popular way in Political Economy to account for fixed effects is a simple

within group-transformation (Baltagi, 2001) that has also been employed by Pitlik and Wirth

(2003).2 This procedure however makes it unfeasible to estimate effects of time-invariant

political and socio-economic variables on reform policies. As has been stated above, some of

the explanatory variables (polity, polconv) of theoretical interest are almost time-invariant for

a number of countries. Moreover, a fixed effects-technique cannot account for level effects of

explanatory variables. It is impossible to distinguish between a country that observed a

continuously maximum level of, say, democracy (e.g. Denmark) and a nation that showed no

democracy at all throughout the entire period (e.g. Syria) by construction. Cross-country

information on political variables but also on economic controls is therefore seriously

disregarded.

A further problem is the dynamic character of the specification, as the dependent

variable lib, which is defined as ∆ efw, is regressed on pre-period efw. Inclusion of a lagged

19

dependent variable (LDV) in the presence of unit fixed effects leads to endogeneity bias, as

the LDV is correlated with the unit effects (Nickell, 1981). Bias is more severe if panels are

short. Endogeneity may also be a problem in the case of political variables (polity, polconv,

checks, govchg). Both economic reform lib and political variables may be driven by a

common, but unknown factor. The estimator provided by Arellano and Bond (1991) deals

both with unit fixed effects and endogeneity by first-differencing the equation and employing

instrumental variables. Yet, first-differencing also has its cost in terms of lost time-series

information and by focusing exclusively on short run-aspects. Any longer run-influences of

political structures are completely disregarded. Furthermore, flaws of a within-estimator in

estimating cross-country level-effects are not cured by the Arellano-Bond-procedure. Hence,

to get additional cross-country and longer run information, a simple pooled OLS-regression

without fixed unit effects may be desirable to make the picture more complete.

The empirical procedure hence is as follows. In section 4.3.1 results for the within-

estimator are reported. In Section 4.3.2 the Arellano and Bond-estimator is employed, and

section 4.3.3 concludes by showing results of pooled OLS-regressions, corrected for

clustering. To control for heteroscedasticity, reported inference in all estimations is based on

robust standard errors.

4.3 Results

4.3.1 Within group-estimator

Table 1 reports results for the within group-estimator. The lagged efw-index has the expected

negative sign and is highly significant in all equations. This shows that there is a general

tendency of countries with a higher level of economic freedom to liberalize less, i.e. a kind of

conditional convergence effect. With respect to the lagged crisis indicators inflation and

growth, only inflation shows the expected relation to economic policy reform. A bad inflation

performance in the previous period is significantly related to more intense reforms in the

20

following period. The growth variable, however, also shows a positive sign, indicating that a

better growth performance in the pre-period is good for policy reform, although it is never

significant at conventional levels. This stands in stark contrast to the crisis-hypothesis. The

result also contradicts Pitlik and Wirth (2003) who find a negative correlation between pre-

period growth performance and economic liberalization. In part, this effect is owed to the

inclusion of time dummies in the present investigation. Results also seem to depend on the

definition of a growth crisis. Economic controls open and secenrol show no relation to lib.

Table 1: Results of within-group estimator

Dependent variable: Economic Liberalization (lib)

(1) (2) (3) (4) (5) (6)

exrestr-variable: polity polity polconv polconv checks checks

efw (t-1) -0.54 -0.55 -0.59 -0.60 -0.57 -0.58 (0.000) (0.000) (0.000) (0.000) (0.000) (0.000)

exrestr 0.22 0.39 0.69 0.74 0.13 0.18 (0.197) (0.029) (0.000) (0.000) (0.161) (0.047)

govchg -0.82 -0.78 -0.81 (0.000) (0.000) (0.000)

inflation (t-1) 0.60 0.58 0.34 0.35 0.38 0.39 (0.000) (0.000) (0.066) (0.052) (0.053) (0.040)

growth (t-1) 1.65 1.54 1.80 1.60 1.85 1.66 (0.156) (0.182) (0.130) (0.169) (0.118) (0.158)

open (t-1) 0.06 0.06 0.01 0.00 0.13 0.14 (0.827) (0.834) (0.975) (0.986) (0.617) (0.602)

secenrol (t-1) -0.44 -0.59 -0.37 -0.59 -0.26 -0.45 (0.483) (0.324) (0.539) (0.311) (0.673) (0.456)

Observations 488 488 493 493 508 508

Number of id 112 112 113 113 117 117

R-squared 0.413 0.438 0.412 0.436 0.389 0.414

F 25.72 25.82 25.90 25.95 24.22 24.44 Note: Period dummies and constant included, but not reported. P-values in parentheses are based on robust standard errors.

21

The behavior of the several indicators of executive restrictions is as expected, as they show

positive signs in all specifications. The democracy indicator polity (columns (1) and (2)) is

however only significant if it is controlled for the average number of government changes

govchg. This indicates that democracy is only good for reform if government is reasonable

stable. This is in line with theoretical assertions on the effects of policy credibility on

economic reforms. These results are fully confirmed using an alternative indicator of political

liberties from Freedom House (2002) (results not reported). The Henisz-indicator of executive

restrictions polconv, shown in columns (3) and (4), is always positively related to

liberalization at a 1 per cent level of significance.3 A third indicator of executive restrictions,

checks, is statistically significant only when it is controlled for govchg, too. Yet, this effect

depends on inclusion of secenrol. If secenrol is eliminated from the set of regressors, checks

also becomes significantly associated with lib in all specifications.4

The indicator of government stability govchg shows the expected highly significant

negative correlation with lib in all specifications. If on average a country's executive is

dismissed once a year, reform effort lib is reduced by 0.8 efw-points. This may yet also be a

result of reverse causation. Thus, all specifications have been re-estimated by employing the

political variables lagged one period. With respect to the several indicators for executive

constraints, the results are fully confirmed with this alternative specifications. However,

lagged govchg completely loses significance. Thus, obtained results may be caused by a

reverse causality. Yet, it may also be the case that only a current executive instability reduces

credibility of reform policies. In the next sub-section, endogeneity is addressed by an

instrumental variable estimator.

22

4.3.2 Arellano and Bond-estimator

Table 2 shows results for the One Step-Arellano and Bond-estimator. As can be seen, first

differencing and instrumenting causes a significant loss of observations. Tests for over-

identification and for (absent) second-order autocorrelation show validity of specifications.

Table 2: Results of Arellano and Bond-estimator

Dependent variable: Economic Liberalization (lib)

(1) (2) (3) (4) (5) (6)

exrestr-variable: polity polity polconv polconv checks checks

efw (t-1) -0.35 -0.36 -0.39 -0.40 -0.31 -0.31 (0.000) (0.000) (0.000) (0.000) (0.000) (0.000)

exrestr 0.42 0.64 1.10 1.18 0.46 0.49 (0.153) (0.047) (0.001) (0.000) (0.003) (0.002)

govchg -1.51 -0.73 -1.32 (0.004) (0.131) (0.007)

inflation (t-1) 0.74 0.75 0.42 0.44 0.46 0.54 (0.001) (0.001) (0.092) (0.054) (0.087) (0.032)

growth (t-1) 1.41 1.22 2.82 2.38 2.85 2.21 (0.581) (0.623) (0.327) (0.384) (0.365) (0.467)

open (t-1) 0.05 -0.13 -0.23 -0.38 -0.11 -0.30 (0.882) (0.756) (0.508) (0.306) (0.767) (0.487)

secenrol (t-1) -1.34 -1.13 -0.38 -0.22 -0.06 0.01 (0.111) (0.146) (0.717) (0.825) (0.951) (0.991)

Observations 375 375 379 379 389 389

Number of id 107 107 108 108 112 112

Sargan (P-Value) (0.322) (0.758) (0.432) (0.437) (0.571) (0.728)

AR (2) (P-Value) (0.655) (0.357) (0.680) (0.535) (0.651) (0.429) Note: Period dummies and constant included, but not reported. P-values in parentheses are based on robust standard errors. Results of Sargan-Test are from a Two Step-Estimator. All explanatory variables are treated as endogenous.

Throughout all models, initial efw-values are again significantly related to lib. Coefficients

however drop from about –0.6 to about –0.35 in first difference estimations. The behavior of

23

all controls and all explanatory political variables is almost similar to the within-estimates.

Neither growth, nor open and secenrol seem to be related to lib. All equations, except for (1),

show a highly significant positive relation between the respective indicators of executive

constraints and the dependent variable. Results for polconv are particularly impressing.

According to estimates (3) and (4), increasing political constraints from 0, which is a typical

value for many lower developed countries, to 0.9 (the common value for Belgium) leads to a

higher liberalization intensity of about 1 point. In two out of three specifications govchg is

negatively related to lib at a 1 per cent-level of significance. In model (4), the P-value is

slightly reduced to 0.13. Thus, even for govchg there is not too much evidence that previous

results from a within-estimator are mainly caused by endogeneity.

4.3.3 Pooled OLS-estimator

Results for a simple pooled OLS-estimator are shown in table 3. In contrast to previous

regressions country dummies are not included here in order to emphasize cross sectional

variation. P-Values reported are based on clustered standard errors, i.e., it is assumed that

observations are independent across countries but not necessarily independent within groups.5

It should be noted first that outcomes in principle confirm results obtained above, where

unit fixed effects were included. Except for lagged growth, coefficients of all variables show

expected signs and are significant at least at a 10 per cent confidence level. The entire set of

covariates indicating restrictions on the executive shows a behavior similar to fixed effects

regressions, though coefficients are generally smaller. Government instability govchg again is

negatively related to economic liberalization lib. Once more, bad inflation performance in the

preceding period leads to higher reform efforts in the current period.

With respect to international trade openness open and to the education-variable secenrol

the level-effects suppressing consequences of including country dummies are seen most

clearly. Both open and secenrol are now highly significant with the expected positive signs. In

24

a cross-country dimension more open economies and economies with a better educated people

observe more intense economic policy reforms, ceteris paribus. In both cases level effects

seem to have been hidden by unit specific effects.

Table 3: Results of Pooled OLS-estimator, no unit fixed effects

Dependent variable: Economic Liberalization (lib)

(1) (2) (3) (4) (5) (6)

exrestr-variable: polity polity polconv polconv checks checks

efw (t-1) -0.18 -0.19 -0.21 -0.22 -0.18 -0.19 (0.000) (0.000) (0.000) (0.000) (0.000) (0.000)

exrestr 0.29 0.38 0.43 0.48 0.13 0.16 (0.002) (0.000) (0.000) (0.000) (0.014) (0.004)

govchg -0.38 -0.29 -0.29 (0.011) (0.042) (0.042)

inflation (t-1) 0.40 0.39 0.26 0.25 0.28 0.28 (0.002) (0.003) (0.085) (0.091) (0.061) (0.067)

growth (t-1) 0.16 0.16 0.13 0.10 0.07 0.03 (0.862) (0.856) (0.886) (0.913) (0.938) (0.970)

open (t-1) 0.16 0.14 0.16 0.14 0.13 0.11 (0.001) (0.004) (0.000) (0.002) (0.003) (0.019)

secenrol (t-1) 0.40 0.39 0.40 0.39 0.49 0.48 (0.007) (0.011) (0.018) (0.020) (0.001) (0.002)

Observations 488 488 493 493 508 508

Number of id 112 112 113 113 117 117

R-squared 0.281 0.289 0.267 0.272 0.250 0.255

F 17.32 15.73 15.81 14.71 15.07 13.92 Note: Period dummies and constant included, but not reported. No country dummies included. P-values in parentheses are based on clustered standard errors.

5 Conclusion

Economic policy reform is a conflict-ridden political process. Beneficial policies for society

as a whole are often not implemented due to a fierce opposition from politically powerful

25

prospective losers of reforms. Thus, it is commonly suspected that the introduction of market-

oriented reforms requires a strong and autonomous government. Institutional restrictions on

the executive appear to weaken decisiveness of state action and constitute a major obstacle to

change. Empirical investigations however often find no evidence supporting the hypothesis of

a superiority of unconstrained governments in executing reforms, so far.

The present paper has reviewed shortly some important contributions to Political

Economy of Reform and attempted at a re-interpretation in the light of credibility problems.

Seen from this view, resistance to beneficial reforms often stems from a lack of government

credibility to keep compensation promises and to not withdraw from a policy change. As

credibility is increasing in the number of procedural restrictions (democracy) and political

veto players, there may be a trade off between decisiveness and resoluteness of a polity.

Moreover, standard theory also disregards that conflict over reform cannot be 'solved' by

unilateral action of an unconstrained executive. In political systems which require a consent

of societal groups for a policy change, distributional conflict is necessarily settled by

negotiations when it comes to an agreement. In an inter-temporal perspective, checks and

balances reduce the risk of an open-ended political dispute and may thereby increase

decisiveness of governmental action.

In the final part of the paper previous empirical results have been put to the test. Two

major problems that have not yet been addressed are (1) that an endogeneity bias due to a

dynamic model specification may determine results, and (2) that the use of a model

specification with country unit effects may eliminate too much cross sectional variation in the

data, especially in the case of almost time invariant political variables. As no method qualifies

as the single best way to estimate the effects of political structures on observed behavior, to

increase confidence in results one might be best advised to compare results from different

estimation techniques as a second best-solution. Doing this, ample evidence has been found in

26

favor of the hypothesis that more constrained governments are also more successful in

executing market-oriented policy changes.

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30

Appendix: Summary statistics

Variable Obs Mean Std. Dev. Min Max

checks 594 0.820 0.635 0 2.625

efw 605 5.362 1.396 1.68 9.03

govchg 541 0.175 0.165 0 0.800

growth 534 0.032 0.035 -0.141 0.146

inflation 584 0.175 0.330 -0.120 3.437

lib 605 0.250 0.664 -2.32 2.65

open 598 0.666 0.479 0.037 3.803

polconv 578 0.384 0.333 0 0.893

polity 573 0.614 0.370 0 1

secenrol 583 0.414 0.205 0.022 0.860

31

Appendix: Correlation matrix (P-Values in parentheses)

lib efw polity polconv checks govchg inflation growth open

efw -0.269 1.000 (t-1) (0.0000)

polity 0.108 0.467 1.000 (0.0095) (0.0000)

polconv 0.072 0.592 0.867 1.000 (0.0833) (0.0000) (0.0000)

checks 0.103 0.435 0.843 0.779 1.000 (0.0121) (0.0000) (0.0000) (0.0000)

govchg -0.006 0.050 0.427 0.300 0.360 1.000 (0.8950) (0.2428) (0.0000) (0.0000) (0.0000)

inflation 0.319 -0.337 0.053 -0.010 0.050 0.016 1.000 (t-1) (0.0000) (0.0000) (0.2158) (0.8209) (0.2283) (0.7091)

growth -0.185 0.229 -0.099 -0.027 -0.062 -0.039 -0.378 1.000 (t-1) (0.0000) (0.0000) (0.0259) (0.5490) (0.1540) (0.3706) (0.0000)

open 0.024 0.405 0.051 0.122 0.081 -0.168 -0.134 0.178 1.000 (t-1) (0.5586) (0.0000) (0.2205) (0.0033) (0.0497) (0.0001) (0.0012) (0.0000)

secenrol 0.057 0.510 0.545 0.605 0.482 0.119 -0.046 -0.063 0.272 (t-1) (0.1731) (0.0000) (0.0000) (0.0000) (0.0000) (0.0067) (0.2750) (0.1546) (0.0000)

32

Endnotes

1The problem may also be reversed, i.e., the expected value of a representative group member is positive and

therefore the entire group supports a policy change. If ex post a majority of the population benefits from the

reform, no damage is done. If ex post a majority of voters are disappointed by the newly enacted policy, it will

be overturned.

2 Pitlik and Wirth (2003) however do not control for time specific effects. Hausman tests clearly reject the use

of a random effects-estimator.

3 All results for polconv in all regressions are confirmed using instead an alternative indicator (polconiii) from

Henisz (2000). In contrast to polconv, the construction of polconiii is based on only three potential veto points.

The judiciary and a decentralized structure are not taken into account in polconiii.

4 However, it is not the other way round. Eliminating indicators of political constraints does not lead to a

significantly positive coefficient of secenrol.

5 I also experimented with panel-corrected standard errors proposed by Beck and Katz (1995). Inferences

however do change only slightly.