National Tax Association is collaborating with JSTOR to digitize, preserve and extend access to National Tax Journal.http://www.jstor.orgNational Tax AssociationAlcohol-Leisure Complementarity: Empirical Estimates and Implications for Tax Policy Author(s): Sarah E. West and Ian W. H. Parry Source:National Tax Journal, Vol. 62, No. 4 (December, 2009), pp. 611-633Published by:National Tax AssociationStable URL:http://www.jstor.org/stable/41790639Accessed: 26-07-2015 20:28 UTCYour use of the JSTOR archive indicates your acceptance of the Terms & Conditions of Use, available at http://www.jstor.org/page/info/about/policies/terms.jspJSTOR is a not-for-profit service that helps scholars, researchers, and students discover, use, and build upon a wide range of content in a trusted digital archive. We use information technology and tools to increase productivity and facilitate new forms of scholarship. For more information about JSTOR, please contact support@jstor.org.This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy Sarah E. West Departmentof Economics , Macalester College , St. Paul MN 55105 Ian W.H. Parry Resources for the Future , Washington , DC 20036 NationalTaxJournal Vol. LXII, No.4 December2009 Abstract - This paperprovides a firstattempt to estimatethe cross-priceelasticity betweenalcoholic beverages andleisure , which is critical forassessing howmuchalcoholtaxation might bewar- rantedon fiscalgrounds. Weestimatea demand systemdefined over alcohol , leisure , andother goods , using data from theConsumer ExpenditureSurvey andothersources.Our results suggest that alcoholisa relative complementfor leisureovera rangeofspecifica- tions.This implies thatthe optimal alcoholtax maysubstantially exceedthe Pigouvian tax , reinforcing the efficiency case forhigher taxation.These findings shouldbeviewedas preliminary , however, given dataandotherlimitations. I. INTRODUCTION Therealcohol are taxes. two potentially First, they important help to mitigate economic various rationales external for alcoholtaxes. First, theyhelp to mitigate variousexternal costsofalcohol abuse, suchas therisksto others posed by drunkdriversandtheburdenofmedicalcostsonthird parties fromalcohol-relatedillness. Second, byraising revenue, they reducetheneedto rely on other taxes, particularly thosethat fallon labor income, tofinancethe government'sbudget. The Ramseytheory of optimal excisetaxation implies thatsucha taxsubstitutionis likely tobe desirableifalcoholis a relative complement for leisure, in whichcase it provides whatwe shalldescribeas an additional"fiscal"rationaleforalcohol taxation.A truly"optimal" alcoholtaxwouldreflectbothof theseconsiderations. As regards thefirst rationale, a numberof studieshave measuredthe Pigouvian, or externality-correcting, level of alcoholtaxes. Although theusual caveatsabout parameter uncertaintyapply, a typical estimateis a taxofaround $70 per gallon, oraboutthreetimescurrentfederaland statetaxesof $24 pergallon; see Parry, West, and Laxminarayan(2009) for a recentdiscussionofthisevidence. Thesecondrationaleforalcoholtaxeshasnotreceived any attentioninthe prior literature, atleastina quantitative sense, so thereis no basis for gauging to whatextentthe optimal alcoholtax may differfromthe Pigouvian tax.In particular, therehave beenno attempts to estimatethealcohol-leisure cross-priceelasticity, even though itis a critical parameter in determining the Ramseycomponent ofthe optimal alcohol 611 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsNATIONALTAXJOURNAL tax.Itbehooveseconomiststo try to esti- matethis elasticity, andthus provide some insight on whetherthereis a valid fiscal rationaleforalcoholtaxes. This paperprovides a first attempt at estimating this elasticity, based on an AlmostIdeal Demand System defined over alcohol,leisure, and other goods, estimatedwithdata fromtheConsumer ExpenditureSurvey and othersources. Whilethedataare probably as good as we can find, and we believewe have pushed the methodology as faras possible with standardeconometric techniques, the confidenceintervalswe obtainon the alcohol-leisure cross-priceelasticity are nonetheless large. However, overa wide range ofmodel specifications, alcoholis a relative complement forleisureovera 95 percent confidence interval, and the resultingRamsey taxis potentiallylarge. In fact, itis quiteplausible thatthefiscal component of the optimal alcoholtaxis more important thanthe Pigouvian com- ponent. Thus, the adjustments foralcohol- leisure complementarity servetoreinforce the efficiency casefor higher alcohol taxes, perhapsby a substantialamount. Given a numberof data and other limitationsdiscussed below, our findings should be viewed as suggestive rather thanconclusive. Nonetheless, our analysis has value in demonstrating the potential empiricalimportance ofthe Ramsey tax, and in developing an econometricmeth- odology thatcanbe refinedas the quantity and quality ofdata improve overtime. Before outlining our empiricalapproach and results, we firstdescribein more detailhow our paper relatesto the pre- vious literatureon optimalcommodity taxes, external costs, and the workplace productivity effectsof alcohol abuse. We concludewitha discussionof some caveatsto thecase for higher taxes, such as equity issues, theriskthatalcoholtax revenueswill not be used judiciously, and politicalopposition to alcohol tax reform. II. BACKGROUND A. Relationtothe Optimal Tax Literature According to Ramsey tax theory, the optimal taxon a commoditymay exceed any level warranted on externality grounds ifthe product is a relatively weak substituteor a strongcomplement for leisure, compared withother consump- tion goods(Sandmo,1975; Bovenberg and Goulder,2002). Underthese conditions, up toa point a revenue-neutralshiftfrom a taxon laborincometo the commodity tax will (slightly) increaselabor supply, inducing an efficiencygain in the tax- distortedlabor market, inadditionto any efficiencygain obtainedfrom mitigating the externality. Converseresults apply if the product is a relativelystrong substitute forleisure. This fiscal component of the optimal commodity tax implicitly combinestwo linkages withthebroadertax system that have been decomposed in theliterature on environmentaltax shifts (Goulder, 1995; Parry and Oates,2000). Firstis the "revenue-recycling effect"or efficiency gain from using additional commodity tax revenues to reduce distortionary labortaxes.Secondis the"tax-interaction effect"or potentialefficiency loss in the labormarketfromthe impact of higher commodityprices on reducing the real returnsto work effort, thereby discour- aging labor supply. For commodities thatarerelativeleisure complements, the revenue-recycling effectcandominatethe tax-interactioneffect (and thelatter may evenreverse sign),implying a netwelfare gain frominteractionswiththetax system, and a positive fiscal component to the optimalcommodity tax. Perhapssurprisingly, therehave been few attempts to apply empirically the optimal commodity tax framework to actual taxes. This may be because economistswere initially concernedwith optimizing over all commodity taxes 612 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy simultaneously, whichis an especially formidable challenge.1 Insteadourfocusis on only one commodity tax, assuming the restof thetax system,primarily income and payroll taxes, is collapsed intoa single tax on labor income (interactions with tobaccotaxesarediscussedinSection V). Even forthe limitednumberof com- modities traditionallytargeted with excisetaxes - primarily tobacco products, transportation fuels, alcoholic beverages, and telecommunications - therehasbeen little attempt toestimate econometrically theleisure cross-price elasticitiesthatare required to assesshow muchtaxationof thesecommodities might be warranted on optimal taxation grounds.2 Diewert and Lawrence (1996) estimate these cross-price elasticitiesformotor vehicles, housing, andother goods, butnotalcohol. Their point estimatesindicatethatboth motorvehiclesand housing are relative leisure complements, with compensated cross-price elasticitiesof-0.14 and -0.79 respectively,compared to a cross-price elasticity of 0.38 for generalconsump- tionwith respect to the price of leisure. Madden (1995) includesbothleisureand alcohol in a demand system estimated with aggregate timeseriesdata forIre- land, though the cross-price elasticities arenot reported. Westand Williams (2007) estimatean AlmostIdeal Demand System defined over gasoline, leisure, and other goods using householddata. They findthatthe optimalgasoline taxis around50 percent larger than the externality-mitigating tax, because gasoline is a relativeleisure complement. Thismakessense intuitively, as gasoline use increasesless than pro- portionately to labor supply(following a compensatedwage increase),given the substantialshare of non-work-related trips in totalhouseholdtravel.Another explanation is that driving is a relatively time-intensive activity atthe margin, once householdshaveincurredthefixedcostof buying a car.Becker's (1965) modeloftime use suggests thattime-intensive goods are complements to non-markettime. Wehavelessintuitionaboutothercross- price elasticities, including thatbetween alcoholand leisure.Howeverwe might expect thatalcoholand leisurewouldbe relative complements, sincea substantial shareof alcoholuse is not work-related, and prior evidence (discussedbelow)sug- gests that spending on alcoholincreases proportionately less than spending on other goods whenhouseholdshavemore (labor) income. B. Relationto theHealthEconomics Literature In health economics, discussionsof optimal alcohol taxes usually focuson the externality-correcting or Pigouvian tax, leaving aside linkages betweenthis tax and the broaderfiscal system. The most importantcomponent of the Pig- ouviantaxis the (marginal) externalcost imposedby drunkdrivers.Thishas been calculatedfromthe (fatal and non-fatal) injury riskthatdrunkdrivers pose tooth- ers, treating risksto occupants invehicles driven by thedrunkdriveras internal.A portion of variousotheraccident costs, such as propertydamage and medical 1 Thiswould require reliableestimatesofown-andleisure-cross price elasticitiesfor every taxed commodity andfunctionalform assumptions that might beunreliablefor largepricechanges. One response tothese problems hasbeentolimitthefocustobroad commoditygroups, andtothe appropriate directionof partial taxreforms (Ahmad andStern 1984;Madden,1995;Zodrow,2006). Anumberofstudiesincludealcoholwhen estimatingcommodity demand systems butdonotinclude leisure, includingAndrikopoulos, Brox, andCarvahlo (1997), BlakeandNied (1997),Browning(1991),Blundell, Pa- shardes, andWeber (1993), DeatonandMuellbauer (1980), Eakinsand Gallagher(2003), FussandWaverman (1987),Gao,Wailes, andCramer (1995), Holm (1995),Jones(1989), NelsonandMoran (1995),Taube,Huth, andMacDonald (1990), andTaubeandMacDonald (1991). 613 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsNATIONALTAXJOURNAL burdens, are also externalto theextent they are borne by insurance companies or the government. Another component ofexternalcostsis themedicalburdenon third parties fromalcohol-inducedillness overthe lifecycle(net of any medicalsav- ings from prematuremortality),though thesecosts appear tobe minorrelativeto thedrunkdriver externality. In comput- ing the Pigouvian tax, roadaccidentand illnesscostsare scaled by the sensitivity of drunk driving and heavy drinking, respectively, to alcohol prices, relative to the pricesensitivity ofoverallalcohol consumption. Studies typicallyput the optimalPigouvian taxat around30 per- centor moreofthe pre-tax alcohol price (Manning et al.,1989,1991; Phelps, 1988; Pogue and Sgontz, 1989; Parry, West, and Laxminarayan,2009). 3 Thereis also a largeempirical literature onthe own-priceelasticity foralcoholcon- sumption,though therearesomeserious measurement challenges in estimating this elasticity(Cook and Moore,2000). Mostestimatesofthe own-priceelasticity forall alcoholic beverages combinedlie betweenabout-0.4 and -1.0 (elasticities forindividual beverages sometimeslie outsidethis range).4 The own-price elas- ticity entersintothe Ramseycomponent ofthe optimal alcoholtax (seebelow): the moreinelasticthe demand, the greater the Ramsey tax (assuming alcoholis a relative leisure complement), as thereis less ero- sionofthealcoholtaxbase in response to higher taxrates.5 Economistshave also estimatedthe productivity and labor supply effectsof alcohol consumption,attributingany negative associationto the inability of heavy drinkersto concentrateon the job or findand retainstable employment (Cook and Moore,2000). As discussed below, such a negative alcohol-health- productivity relation plays a separate and reinforcing rolein raising the optimal alcoholtaxabovethe Pigouvian tax.Some studiesfinda negative association (Mul- lahy and Sindelar,1991,1993), butothers finda "drinker's bonus/' that is, a positive associationbetween earnings and alcohol consumption(Berger and Leigh, 1988; Zarkinet al., 1998)though theseresults may be biased, as earnings area determi- nantofalcohol consumption. Tworecentstudiesaddressthe possibil- ity of endogeneitybyestimating reduced formmodels relating labormarketout- comesto alcoholtaxes.WhileDave and Kaestner (2002) findno evidence that alcoholtaxesaffect wages,employment, or hours, theyexpress concernthat specifi- cationerror may obscurethetruerelation- ships. Cook and Peters (2005)attempt to avoidthis problembyusinglongitudinal data, whichenablesthemto controlfor a large set of individual-specific charac- teristics.Theirresultsshow a positive relationship betweenalcohol taxesand earnings and support thenotionthatthe drinker'sbonus found by othersis due to reverse causality,given thatalcoholis a normal good. Because our study focuses on the sensitivity of alcohol consumption with respect tothe price ofleisure (and notthe reverse), we avoidthekindof endogeneity problem discussed above, butfaceanother ifalcohol consumption affects wages. We use instrumentalvariable techniques to addressthis problem, as discussedbelow. 3 Kenkel (1996) estimatesa somewhat larger taxwhen peoplemisperceive the long-term risksfrom heavy drinking. 4 Somerecentstudiesinclude Baltagi andGoel (1990),Baltagi andGriffin (1995), Leeand Tremblay(1992), Man- ning,Blumberg, andMoulton (1995),Manning and Mullahy(1998), NelsonandMoran (1995), Selvanathan (1991), andYen (1994). 5 Ontheother hand, a moreinelasticdemandreducesthe externality benefitsfroma given alcoholtax (though itdoesnotaffectthe Pigouvian tax, ifthedrunkdriverand heavydrinking elasticitiesarealso proportionately smaller). 614 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy III. ECONOMETRICMETHODOLOGY A. Model Specification We specify the following AlmostIdeal Demand System(AIDS) foran individual household h, definedoverthree "goods", alcohol (A), leisure (/), and a composite of all other consumptiongoods (C):6 (la) s'; = a1; + Jjniklogp'' k +j / PhX j, k = alcohol, leisure , other goods (lb) logP>=Is*log(p,7p;.) i (c) 5>*=i,2X=0'5^=0, iii jk = kj (Id) a,*=?, + I/X wheres h isthe expenditure sharefor good;, Fhis fullincome (total spending on alcohol, leisure, andother goods), pkh is the price of good k faced by household h, and Phis a unitinvariant price index, where p. is the mean price for goodj overall households. Thecoefficients ah, jujk, and j are param- etersto be estimated, after imposing the restrictionsin (lc), whichfollowbecause household budget sharessum to unity, and demandfunctionsare homogeneous of degree zero in prices and fullincome and satisfySlutskysymmetry.7 In (Id), we allowa vectorofhousehold specific characteristics (age, race, educa- tion,etc.), indexed by r' toaffect demand, where jQ and are parameters tobe esti- matedand e!1is an errorterm reflecting unobserveddifferencesin preferences. We alsoincludestatefixedeffectsand regional dummies (Northeast,Midwest, Southand West) intoaccountforunobservedlocal factorsthat might affectalcohol use or workbehavior (e.g.,liquor laws, cultural factors,climate, and job opportunities). As discussed below, the Ramsey taxor fiscal component of the optimal alcohol tax depends on the compensated elastic- ity ofalcoholdemandwith respect tothe price of leisure, whichis themainfocus ofourestimation. B. Data Sources 1. HouseholdData Our main data set consistsof 9,454 householdobservationsfromtwelvecon- secutive quarters oftheConsumer Expen- diture Surveys(CEX), 1996-1998.Each quarter, 20 percent ofthe sample is rotated outand replacedby new households; we pool householdobservationsacrossdif- ferent quarters.8 The CEX files include, foreach household, spending on alcohol, total spending on all goods, numberof children, and stateofresidence.9Another expenditure filein theCEX containstwo categories of spending on alcohol: (1) 6 TheAIDS provides a first-order approximation to any demand system andsatisfiestheaxiomsofconsumer choice (Deaton andMuellbauer 1980). Unlikecertainotherdemand systems, itdoesnot impose(1) weak separability betweenleisureand consumptiongoods, or (2) homothetic preferences(which implyunitary expenditure elasticitiesforall goods); eitheroftheserestrictionscould seriously biasestimatesofthealcohol- leisure cross-priceelasticity. 7 Welimitthenumberof goods tothreefor tractability. Theomissionofanother specificgood wouldbiasthe estimateofour quantity of interest,thealcohol-leisure cross-priceelasticity, iftheeffectofthe wage onalcohol consumption iscorrelatedwiththeeffectofthepriceoftheotherspecificer>ndnnalrnhnlmnQiimntinn 8 Pooling observationsenablesusto exploit variationovertimewithinhousehold.Asmentioned below,we cluster by householdtocorrectfor any biasdueto dependenceacrosstheseobservations. 9 Self-reported alcohol consumptionusually understatesactual consumption,perhapsby asmuchas50 percent (Cook and Moore,2000). If underreporting varieswiththe wage, thenourestimatesofthe cross-price elastic- ity ofdemandwillbebiased.For example, more prevalent underreportingamonghigh-wage workerswill biasthe cross-price estimatedownward. However,thereis very littleevidenceonwhethertheerrorsacross surveyrespondents are additive, proportional, randomor systematic, soitisunclear whether, andinwhat way,underreporting affectsourresults. 615 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsNATIONALTAXJOURNAL beerand wine, and (2) all otheralcohol (i.e., spirits). We attempted to estimate separate elasticitiesforeach category but resultswere implausible or imprecise, as many households only consumeone of thetwo categories.10 In addition, foreach household, thefilesincludeusual weekly work hours, occupation, the gross amount of last pay, the durationof thelast pay period,age, race,sex, and educationlevel (which we codeas above, equal, orbelow high school diploma).11 We calculate (weekly)spending on the compositegood as total expenditure less thaton alcohol, leisureas a (non-sleep) time endowmentof90hours per weeklesswork hours (our resultsare notsensitivewith respect to thevalue of thetimeendow- ment), andfullincomeas total expenditure plus the product of leisureand the net wage.Wages aremeasured bygrosswages fromthe CEX, correctedforselectionbias (see below), and netof federaland state incometaxesandearnedincomeandchild tax credits; effectivetaxratesfordifferent wage rateswereobtainedfromNBER's TAXSIMmodel (Feenberg and Coutts, 1993). As payroll taxesare partially offset byhigher futuresocial security benefits, we do notdeducttheminourbaseline specifi- cation, though we do deductthemin one ofthe sensitivityanalysisspecifications. 2. PriceData Forthealcohol price we use theAmeri- can Chamber of CommerceResearch Association (ACCRA) cost-of-living index, whichlists quarterlyaverageprices for beer,wine, and spirits for approxi- mately 300 urbanareas.We weightcity and town pricesby population shares (from the 2000 census) to obtainstate- levelalcohol prices. Toobtainone price of alcohol, we convert beer,wine, and spirits prices into pricesper liter, and weight them by the average shareofeachbever- age intotallitersofalcohol consumption. TheACCRAdataarealsousedtoobtaina price indexforthe compositegood. Although oftenused in the empirical alcohol demand literature, theACCRA price index can be problematic. One issue is thatthe data are collected by membersoflocalchambersofcommerce and there may be some inconsistency in measurementacross states. However, measurementerrorsin alcohol prices are only a concernforthealcohol-leisure cross-priceelasticity if they are spatially correlatedwith wages or leisure, which seems unlikely. Another problem is that spatial differ- encesin alcohol pricesmay be collinear withthe statefixed-effectvariables.In our data, state fixed effectsabsorb a significantportion but by no meansall ofthevariationin statealcohol prices, as these pricesvary acrosstime.The average standarddeviationsofwithin-state wine, beer, and alcohol prices are 11 percent, 9 percent, and 7 percent ofthewithin-state priceaverages,respectively. Thereis on average a differenceofabout $2,$4, and $8, respectively betweenminimumand maximum prices withinstates. After these prices are weighted and averaged toobtainone priceper literof alcohol, the average within-statestandarddeviationin 10 Assessing whetherthereis any basisfordifferentialtaxationofindividual beverages on externalitygrounds is very difficultbecause, for example, dataonalcoholinvolvementintrafficaccidentsisnot distinguishedby beverage class.Asforthefiscalrationale, there appears tobesomebasisfor taxing beermore heavily than wine, andwinemore heavily than spirits(Parry, West, and Laxminarayan,2009). Thisisbecausetheestimated own-price elasticitiesforbeeraresmallerin magnitude thanforwine, whichinturnaresmallerthanfor spirits. In contrast, spirits aretaxedmost heavily, atabout$35 pergallon ofalcohol, compared with$20 pergallon for beer, and$18forwine (Parry, West, and Laxminarayan, 2009, Table 1). 11 Otherdatasets commonly usedm empirical workonalcoholdemanddonotcontamalltnevariaoieswe require; for example, theNationalHealthInterview Survey excludes wages, whiletheBehavioralRiskFactor Surveillance System excludesboth wages andhoursworked. 616 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy this price is 8 percent ofthe averageprice. Lack ofvariationin within-statealcohol pricesmagnifies the standarderroron the own-price alcohol elasticity,though again thisis not a primary concernfor thealcohol-leisure cross-priceelasticity. Nonetheless, we also estimate specifi- cations (1) withno fixed effects, and (2) withvarious fixedeffectsalternatives commonly used in the literature, includ- ingproportion ofstate population in col- lege, state averagetemperature,average cloudydaysperyear, numberof drinking establishmentsand places of worshipper capita, anda dummy variableforwhether alcoholsales are permitted on Sundays (toproxy for anti-drinkingsentiment). In thesealternative specifications theconfi- denceintervalforthe own-price alcohol elasticity is somewhat narrower, though thatforthe alcohol-leisure cross-price elasticity is onlymoderately affected.In thebenchmarkestimationwe controlfor statefixedeffectstoreducethe possibility ofomittedvariablesbias.12 C. Correcting forSelectionBias and Endogeneity Notall householdmembers participate in thelaborforceand abouthalfof the householdsinour samplereport noalcohol consumption.Following Heckman (1979), we attempt tocorrectfor possible selection bias by estimatingprobits on thediscrete choiceofwhethertoworkandwhetherto consumealcohol.Wethenexcludefromthe second-stage estimationhouseholdsthat do notworkandthatdo notconsumealco- hol, to avoid estimationbias whenthere area large numberofcensoredhouseholds (Shonkwiler and Yen,1999). Moredetails on these first-stage estimation procedures are provided in Appendix A.13 In thecase oflaborforce participation, we estimatea probit model jointly with a wage equationusing thefullinforma- tion maximumlikelihood approach to generate the selectivity-correctedwage foreach householdforuse in estimating the demand system in (la). In thefirst- stage estimationwe use standardexclu- sionrestrictionsfromtheliterature (e.g., numberof children, partner'searnings, and the unemploymentrate). Forthedecisionofwhethertoconsume alcohol, we estimatea probit toobtainthe predicted inverseMills ratios, denoted MRha, and includethemin the demand system in (la) to give: (2) s) = ;j0 + Xc,/+J^fijk logphk rk +t log(F'' /Ph) + y . MR*+ e* where yj are parameters. However, our exclusionrestrictionforalcohol consump- tion, whethertheindividualis overthe age of 21, is probably a weak sourceof identificationas it involves relatively few households, and as beingunderage affectsalcohol consumption on boththe extensiveand intensive margins.14 In the sensitivityanalysis we thereforealso fol- low many otherstudiesin theliterature 12 Someresearchershaveusedstatebeertaxesto proxy foralcohol prices, butthisisalso problematic(Young and Bielinska-Kwapisz,2002). In particular, taxesare only a tiny fractionofretail prices, andthereforefail tocontrolfor spatial differencesin transportation, distribution, andother producer costs.We experimented withstate-and quarter-specific beertaxratesasinstrumentsforalcohol prices, butthisrenderedthealcohol own-priceelasticitypositive. 13 Theusualcaveatabout problems with using theHeckmancorrection applies here - only data resulting from a natural experiment(withperfectmeasurement) would yieldfully crediblesolutionsto censoringproblems. Unfortunately nosuchdataexist. 14 Variablesrelatedto religion are probably more likely toaffect only theextensive margin, andthusbebetter exclusionrestrictions. Unfortunately, theCEXdoesnotaskhouseholdsabout religious affiliation.Itdoesask householdshowmuch theygive to religiousorganizations, butlessthan8 percent ofourhouseholds responded tothe question. 617 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsNATIONALTAXJOURNAL Table1 Summary Statisticsfor Working HouseholdswithPositiveAlcohol Consumption One-AdultHouseholdsTwo-AdultHouseholds StandardStandard VariableMeanDeviation MeanDeviation Weekly alcohol consumption(liters) 2.082.802.112.84 One-adult weekly workhours40.511.4 Two-adultmale weekly workhours44.910.2 Two-adultfemale weekly workhours37.611.3 Shareofalcohol expenditure infullincome.02.03.007.01 One-adultshareofleisureinfullincome.49.15 Two-adultmaleshareofleisureinfullincome.28.08 Two-adultfemaleshareofleisureinfullincome.25.08 Shareof compositegood infullincome.50.16.46.13 Weighed alcohol price($/liter) 5.84.445.83.45 Priceof compositegood(index) 1.04.101.03.11 One-adultHeckman-correctednet hourlywage($) 8.092.42 Two-adultmaleHeckman-correctednet hourlywage($) 10.983.29 Two-adultfemaleHeckman-correctednet hourlywage($) 8.562.20 Weekly fullincome ($) 879.70413.60 1,858.60 728.10 One-adult age(years) 37.311.70 Two-adultmale age(years) 38.49.9 Two-adultfemale age(years) 36.99.4 One-adulteducation:< high school diploma(%) 8.5 One-adulteducation: high school diploma(%) 23.9 One-adulteducation:> high school diploma(%) 67.6 Two-adultmaleeducation:< high school diploma(%) 8.7 Two-adultmaleeducation: high school diploma(%) 27.9 Two-adultmaleeducation:> high school diploma(%) 63.4 Two-adultfemaleeducation:< high school diploma(%) 7.3 Two-adultfemaleeducation: high school diploma(%) 26.3 Two-adultfemaleeducation:> high school diploma(%) 66.3 Numberofchildren.42.881.171.19 Race of householdhead White (%) 87.887.6 Black (%) 14.78.6 Asian (%) .6.8 Otherrace (%) 2.93.0 Region Northeast (%) 15.413.4 Midwest (%) 23.025.4 South (%) 34.035.7 West (%) 27.525.4 Observations3,3953,242 byignoring thediscretechoiceof alcohol, reporting resultsfrom system estimation on thenon-censored sample.15 The net-of-tax wage rate might be endogenous if (1) alcohol abuse affects on-the-jobproductivity, (2) thereare errorsin measuringearnings and hours worked that are correlated, and (3) marginal incometax rates vary as earn- ingsvary withhoursworked.To obtain consistentestimateswe instrumentfor wages in our benchmarkcase using the occupation-, state-, and gender-specific mean net wage, fromthe entireCEX. Because observationsare thin across some quarter-occupation-state-gender categories, we use time-invariant wage instrumentsin ourbaseline specification, 15 Wealsoestimateda two-part modelwithnoexclusionrestriction; butidentificationoftheself-selectionmodel throughnon-linearity oftheinverseMillsratioalone may alsobeweakas thisratioislinearovercertain ranges oftheindex (Velia,1998). Thisestimationalso yields a positive(butinsignificant)own-priceelasticity for alcohol; again thiscastssomedoubtonthe reliability ofourestimateofthis elasticity, butisnota major concernforthealcohol-leisure cross-priceelasticity. 618 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy but we also report resultsfrom using quarter-specificwage instruments.The real incometerm log(Fh/Ph)may also be endogenous, because Phis a functionof individual-specificexpenditure shares thatare also dependent variables.We thereforeinstrumentforthis term, using an alternative price index obtained by replacing the individual-specific shares in equation (lb) withthe sample mean shares.Becausethe instrumentis stilla functionof thesame ratioof household price to mean price in equation (lb), it is strongly correlatedwiththeinstrumented variable, butis no longer a functionofthe dependent variables. D. Summary Statistics Table1 presentssummary statisticsfor data used in our benchmark estimation, that is, for working households with positive alcohol purchases. One-andtwo- adult householdseach consumeabout twolitersofalcohol per week, orabout1-2 percent ofhouseholdfull income, while bothhousehold typesspend abouthalfof theirfullincomeon leisure.The average selectivity-corrected net wage is $8.09 per hourin theone-adult sample, and $10.98 per hourformenand $8.56 per hourfor womenin thetwo-adult sample. E. EstimationProcedure Weuse three-stage least squares(3SLS) to estimate separate demand systems for one- and two-adult households, which enablesus to use instrumentalvariables and generalized least squares to account for any errorcorrelationacross equa- tions.16Each adult's leisure is treated as a separategood; thus, the two-adult demand system includesmale leisure, female leisure,alcohol, and composite consumption. We impose therestrictions in (la-d) and drop the equation forother goods, as the restrictionsconstrainthe parameters fromthat equation tobe linear combinationsoftheestimated parameters in the gasoline and leisure equations. Householdcharacteristicsincludemem- bers' age, age-squared, race, sex (in one- adultestimation only), numberofchildren and educationlevel.17 We use parameters fromthedemand system to generateaggregate alcohol demandelasticitiesforone-andtwo-adult households, and alcohol-leisurecross- price elasticitiesforone-adulthouseholds and formen and women in two-adult households.18To obtaina single, own- price alcohol elasticity, we takea weighted average overthoseforone-andtwo-adult householdswherethe weights are their alcohol consumption. And to obtaina single alcohol-leisure cross-price elastic- ity, we first average themaleand female elasticitiesfortwo-adulthouseholdsand thentake an alcohol-weightedaverage over elasticitiesforone- and two-adult households. Aggregate labor supply elasticitiesarecalculatedina similar way, weightingby hoursworked. Confidenceintervalsforelasticitieswere obtainedfroma non-parametric bias-cor- 16 We experimented with estimating the systemusinggeneralized methodofmoments (GMM). Sincesuch estimationdidnot appreciablychange estimatesnor improveprecision andismoretediousto implement, weelectedtouse3SLS (but also report resultsobtained using2SLS). 17 Wecouldestimatethefulleconometricmodel, including alldiscreteandcontinuouschoices, withmaximum likelihoodestimation.However, since censoring occursinbothalcoholandleisuredemand, andforeither orboththemaleandfemaleintwo-adulthouseholds, wewouldneedtoevaluate multipleintegrals inthe likelihood function, whichwouldbe computationally intensive given thatwe bootstrap standarderrors. 18 Wecalculateelasticities using methodsand equationsexplained inWestandWilliams (2007). Parametersfrom the system estimationareusedtofindtheeffectsof prices on consumption shares, whicharethentransformed intoderivatives involvingquantities, calculatedforeachhouseholdand correcting for any cornersolutions. Finally, these household-specific derivativesaretransformedand aggregated into sample-wide elasticities. 619 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsNATIONALTAXJOURNAL Table2 One-AdultHouseholdDemand System EstimationResults (Baseline) AlcoholShareLeisureShare ln(alcoholprice) -0.0012-0.0066 (0.0335)(0.0062) ln(othergoodprice) 0.0077-0.0808*** (0.0354)(0.0352) ln(netwage) -0.00660.0874*** (0.0062)(0.0343) In (F/P) -0.0152***-0.3504*** (0.0036)(0.0282) Inversemillsratio (alcohol) 0.0088-0.3275 (0.0896)(0.5013) Age -0.0006-0.0037 (0.0006)(0.0042) Agesquared 0.0000070.00004 (0.000007)(0.00005) Black-0.00240.0137 (0.0054)(0.0360) Asian-0.00650.0084 (0.0118)(0.0927) Otherrace-0.00100.0222 (0.0051)(0.0359) High school degree 0.00160.0430* (0.0036)(0.0256) Morethan high school degree 0.00020.0495 (0.0055)(0.0367) Female0.00080.0116 (0.0015)(0.0182) Numberofchildren0.00050.0033 (0.0012)(0.0073) Constant0.1314***2.8864*** (0.0657)(1.467) Numberofobservations3,3953,395 Notes:These3SLS regressions use ln(mean net wagebyoccupation,by stateand gender) instrumentsfor ln(netwage) and ln(F/P) calculated using the price indexbasedonmean expenditure sharesas instru- mentsforthe ln(F/P),usingindividual-specific shares.All regressions includestateand regiondummy variables. Bootstrapped standarderrorsarein parentheses. Asterisksindicate statisticallysignificance atthe 1% (***), or10% (*) level. rected bootstrap thatselects 1,500 random sub-samples ofthefulldata set, estimates the correctionsfor selectivity bias, and thenestimatesthedemand systemsusing each sub-sample. Weclusterobservations by householdin generating each bootstrap sample,given thatobservationsforthe samehouseholdfor multiplequarters are not independent (thisprecludes us from clusteringbyany other variable).19 19 Wealso attempted tocluster by stateratherthan by household, butthisworsensthefitonthe compensated labor supplyelasticity andthereforeonthedifferencebetweenitandthe cross-priceelasticity. Thebias-corrected bootstrap methodis appropriate incaseswherethevariances vary asa functionofthe parameters ofinterest. 620 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy Table3 Two-AdultHouseholdDemand System EstimationResults (Baseline) AlcoholShareMaleLeisureFemaleLeisure Ln(alcoholprice) -0.0021-0.0039***-0.0011 (0.0042)(0.0015)(0.0016) Ln(othergoodprice) 0.0071***-0.067***-0.0464*** (0.0035)(0.0164)(0.0110) Ln(male net wage) -0.0039***0.1455***-0.0746*** (0.0015)(0.0137)(0.0074) Ln(female net wage) -0.0011-0.0746***0.1221*** (0.0016)(0.0074)(0.0087) Ln (F/P) -0.0070***-0.169***-0.1684*** (0.0014)(0.0305)(0.0230) Inversemillsratio (alcohol) -0.00630.28580.2690 (0.0212)(0.3286)(0.2450) Male age 0.0003-0.0043-0.0018 (0.0003)(0.0027)(0.0019) Male agesquared -0.0000040.00005*0.00002 (0.000003)(0.00003)(0.00002) Blackmale-0.0002-0.0170-0.0331 (0.0029)(0.0533)(0.0416) Asianmale-0.0046*0.03040.0066 (0.0025)(0.0445)(0.0352) Otherracemale-0.0005-0.00940.0179 (0.0019)(0.0343)(0.0250) Male high school degree 0.0012-0.0077-0.0075 (0.0017)(0.0214)(0.0170) Malemorethan high school degree 0.0015-0.0091-0.0089 (0.0018)(0.0217)(0.0153) Female age -0.00010.0017-0.0017 (0.0002)(0.0022)(0.0019) Female agesquared 0.000001-0.000010.00003 (0.000003)(0.00003)(0.00002) Blackfemale0.00080.01960.0261 (0.0030)(0.0544)(0.0421) Asianfemale-0.0004-0.033-0.0412 (0.0037)(0.0618)(0.0445) Otherracefemale-0.00010.0205-0.0147 (0.0021)(0.0367)(0.0288) Female high school degree -0.0011-0.00710.0002 (0.0014)(0.0154)(0.0128) Femalemorethan high school degree -0.0019-0.0050.008 (0.0014)(0.0149)(0.0127) Numberofchildren0.0001-0.00070.0074 (0.0002)(0.0036)(0.0028) Constant0.0756*1.1901*1.2493* (0.0432)(0.6499)(0.6684) Observations 3,2423,2423,242 Notes:These3SLS regressions use ln(mean net wagebyoccupation,by stateand gender) instrumentsfor ln(net wage) and ln(F/P) calculated using the price indexbasedonmean expenditure sharesas instrumentsforthe ln (F/P)usingindividual-specific shares.All regressions includestateand regiondummy variables. Bootstrapped standarderrorsarein parentheses. Asterisksindicate statisticallysignificance atthe1% (***), or10% (*) level. 621 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsNATIONALTAXJOURNAL IV. RESULTS A. EconometricEstimates 1. Regression Results Tables2 and3 report thecoefficientesti- matesforthebaseline specification ofone- and two-adultdemand systems, which includesthevariablesdescribedabove. Negative coefficientson the ln(F/P) terms indicatethatforbothone-and two-adult households, alcoholandleisureareneces- sities.In one-adult households, theshare ofleisureincreasesas wage increases.For two-adult households, theshareofleisure (for either adult) also increasesas his or her wage increasesbut decreasesas the wage oftheotheradultin thehousehold increases.20The effectof themale wage on alcohol consumption is the strongest ofthe cross-pricerelationships estimated here; the shareof spending devotedto alcoholfallsas themale wage increases. 2. BaselineElasticities The firstrowofTable4 reports elastic- ity estimatesand 95 percent confidence intervalsforour two-step baseline specifi- cationwithstatefixedeffects.Confidence intervalsforthe uncompensated and compensated labor supply elasticities are 0.02-0.28and 0.19-0.39 respectively, whichareconsistentwith prior estimates inthe empirical microliterature (Blundell and MaCurdy, 1999;Fuchs, Krueger, and Poterba,1998).21 Alcoholis a relative complement for leisurewhenthe (compensated) alcohol- leisure cross-priceelasticity is less than the compensated labor supplyelasticity (see below); itis an absolute complement forleisurewhenthe cross-priceelasticity is negative. In ourbaseline specification, theestimated cross-priceelasticity is -0.09; however, ithasa wideconfidenceinterval of -0.42 to 0.22, underscoring the need for sensitivityanalysis in inferringopti- mal taxes, ratherthan placing too much emphasis on thebaseline point estimate. The lastcolumnofTable4 indicatesthat alcoholis atleasta relative (if not absolute) leisure complement over a 95 percent confidenceinterval. Ourbaselineestimateofthe own-price alcohol elasticity is -1.19, whichis on the high side relativeto earlier literature, though the confidenceintervalforthis elasticity is very wide, reflecting limited spatial variationin alcohol prices.Again, however, our primary focusis noton this elasticity,given thatthereis reasonable consensus among healtheconomistsover a plausiblerange forits magnitude. We also estimatethe aggregateexpenditure elasticity of demandforalcoholat 0.06 (not showninthe table), whichis broadly consistentwith previous studies.22 20 Whentranslatedinto cross-price elasticities using the techniques describedinfootnote17 above, theseeffects becomemoreintuitive.Asshownbelow, wefind positivewage elasticitiesoflabor supply; anincreaseinthe wage causestheshareofleisuretoincreasenotbecausehouseholdsincreaseleisure hours, butbecausethe decreaseinleisurehoursis proportionally smallerthanthe wage increase. Similarly, our parameter estimates translateinto negativecross-price labor supply elasticities; asmen's wage increases, women'sworkhours decrease, andaswomen's wage increases, men'sworkhoursdecrease. 21 They arealso broadly consistentwithlabor supplyassumptions intaxsimulationmodels (browning, 190/; Ballard,Shoven, and Whalley, 1985;Ballard,1990; Goulderand Williams,2003). Even though the uncompen- satedhoursworked elasticity formalesis typically estimatedtobeclosetozeroor slightlynegative, estimates ofthe economy- wide elasticity,averaged overhoursworkedand participationresponses formaleandfemale workers, are generallypositive; this mainly reflectsthesizable participationelasticity for secondary workers. Macroeconomicstudiesthat attempt to explainaggregate labor supply variationacrossbusiness cycles, or acrossdifferent countries, findmuch larger elasticitiesthanthemicrostudiesinthe empirical laborliterature, though thereasonsforthis discrepancy remaina puzzle(Prescott,2004). 22 Recentestimatesof expenditure elasticities (averaging overall beverages) includeU.IUin baitagi andL.nrnn (1995), below0.10in Farrell, Manning, andFinch (2003), 0.11inLeeand Tremblay(1992), 0.25in Manning, Blumberg, andMoulton (1995), 0.40inNelsonandMoran (1995), 0.18inRuhm (1995), 0.89inSelvanathan (1991), and0.4inYen (1994). 622 This content downloaded from 129.15.14.53 on Sun, 26 Jul 2015 20:28:43 UTCAll use subject to JSTOR Terms and ConditionsAlcohol-Leisure Complementarity:Empirical Estimatesand Implications forTax Policy "03 t & c T G U * as W 43 S w - I S .s a I n- R n- vs;s;; s.s.s. 3;s;s;I . od g g o ^ diSdi2ddo d^ o ^ d > "re S ai&< gX ^ 5 I i .y -3 *3oS^ B ^?g8io 'S & n ^ ^^ ^^ ^ ^^ ^ ^ ^^ ^ & " & dH 2^ C U OOO