bUniversity of Oslo, Department of Economics, Moltke Moes vei 31, 0851 Oslo, Norway
a r t i c le i n f o
JEL classification:
H24 H52 J13 J22 Keywords:
Child care Female labor supply Tax revenue Instrumental variables
a b s t r a ct
WestudytheimpactofchildcarefortoddlersonthelaborsupplyofmothersandfathersinNorway.Foridenti- fication,weexploitthestaggeredexpansionacrossmunicipalitiesfollowingalargechildcarereformfrom2002.
OurIV-estimatesindicatethatchildcarecausesanincreaseinthelaborsupplyofcohabitingmotherswhomove towardsfulltimeemployment.Despitethis,averagetaxespaidontheextraincomeislow,lendinglittlesupport totheargumentthatpartsofthecostofchildcareisoffsetbyincreasedtaxes.Meanwhile,wefindnoimpactfor fathers.
1. Introduction
Overthelastdecade,policymakershaveshownincreasinginterest ingovernmentinterventionsinthemarketforchildcare.TheOECDhas discussedtheintroductionof earlychildhoodprogramsin severalre- ports(Fieldetal.,2007;OECD,2006);inGermany,SouthKorea,Canada andtheScandinaviancountries,governmentshavebeenpushingtoex- pandaccesstosubsidizedcare.IntheUS,PresidentObamaproposed tomake“high-qualitypreschoolavailabletoeverychildinAmerica” in hisStateoftheUnionaddress2013.
Animportantargumentinfavorofgovernmentssubsidizingandfa- cilitatingchildcareavailabilityistheclaimthatchildcarehelpsrecon- cileworkandfamilyresponsibilities,therebyincreasingmothers’labor forceparticipation(OECD,2006).Apositiveimpactonlaborsupplymay alsomeanthatthepubliccostofprovidingchildcareispartlymitigated byincreasesinthetaxbaseorreducedbenefitdependence.
Inthispaper,weestimatethelaborsupplyandtaximpactsofchild careusingalargeincreaseinchildcareavailabilityfortoddlers(age2) inNorwayfollowingareformfrom2002toexpandchildcaretocover demand.Thereformincreasedgovernmentsubsidiestoinvestmentin
☆WethankNinaDrange,KjetilTelle,OddbjørnRaaum,KalleMoene,MoniquedeHaan,JoshuaAngrist,BeataJavorcik,RagnhildBalsvik,RagnarTorvik,KaiLiu, ManudeepBhuller,EdwinLeuven,threeanonymousrefereesandseminarparticipantsatUiO,NHH,OFS,Berkeley,Toulouse,TheStavanger-BergenWorkshopon LaborEconomics,TheNorwegianTaxForum,TheNorwegianEconomicsMeetingandtheHarrisSchoolofPublicPolicyattheUniversityofChicagoforhelpful commentsandsuggestions.FinancialsupportfromtheNorwegianResearchCouncil(236947)isgratefullyacknowledged.Theprojectisalsopartoftheresearch activitiesatOsloFiscalStudiesandtheESOPcenterattheDepartmentofEconomics,UniversityofOslo,supportedbyTheResearchCouncilofNorway.Workona previousversionofthisprojectwascarriedoutaspartofproject1172SocialInsuranceandLaborMarketInclusioninNorwayattheRagnarFrischCentreforEconomic Research.
∗Correspondingauthor.
E-mailaddresses:martin.andresen@ssb.no(M.Eckhoff Andresen),tarjei.havnes@econ.uio.no(T.Havnes).
andrunningofchildcareinstitutions,andgeneratedlargevariationin thesupplyofchildcarebetweenmunicipalitiesandovertime.Ouresti- mationstrategyexploitsthedifferenceinchildcareexpansionbetween municipalitiesfollowingthisreformtogetcredibleestimatesonhow child careaffectsthelabor supplyof mothersandfathers,aswell as otherpotentialcaregivers.Becausechildcarefortoddlerswasstrongly rationedinthisperiod,changesintheavailabilityshouldbedrivenpri- marilybythechangesinsupplyasaresultofthereform,andnotby changesin localdemand.Toguardagainstomittedvariablebias,we nonethelessdocumentthatourestimatesarerobusttocontrollingfor alargesetofobservablecharacteristicsandtoarangeofspecification checks.
Animportantimprovementovermuchoftheliteratureisthatweob- serveacontinuousmeasureoftheindividualuseofchildcarethrough- outtheyear.ThisallowsustoimplementanIVstrategy,usingtheavail- abilityofchildcareasaninstrumentfortheactualenrollment,which accountsfortheintensityofchildcareuse.Ourestimatesreflect,there- fore,howafullyearofchildcareenrollmentaffectsmothers’laborsup- ply,ratherthanthedirecteffectofadditionalchildcareslotsthatopenat somepointduringtheyear,whichisestimatedinmostoftheliterature.
https://doi.org/10.1016/j.labeco.2019.101762
Received25May2018;Receivedinrevisedform5September2019;Accepted7September2019 Availableonline9September2019
0927-5371/© 2019TheAuthors.PublishedbyElsevierB.V.ThisisanopenaccessarticleundertheCCBY-NC-NDlicense.
(http://creativecommons.org/licenses/by-nc-nd/4.0/)
Thereareatleasttworeasonswhyaccountingfortheintensityisimpor- tant.First,newchildcareplacesaretypicallycreatedinAugust,atthe startoftheschoolyear.Thismechanicallygivesamaximumchildcare useofaround42%formostchildren,i.e.5outof12months.Second,to thelimitedextentthatchildcareplacesarenotutilizedtocapacity,child careinstitutionshavequiteextensivediscretiontoscaleprovisiontoac- tualuse,bye.g.adjustingemployeehours.Inthiscase,treatingchild careplacesasuniformindependentofutilizationwilloverestimatethe costofprovision.Afurtherimprovementovermuchofthepreviouslit- eratureisthatweidentifybothcohabitingandsinglemothers,aswell asfathersandgrandparents.
Thisisamongthefirstpaperstoprovideevidenceonhowlarge-scale, universalchildcarefortoddlersaffectsparentallaborsupply.1Incon- trast,thesizableliteratureonchildcareandmotherslaborsupplyhas beenfocusedmostlyonchildcareforpreschoolers(age3–6),forwhom thereeareimportantreasonstobelievethattheimpactonparentalla- borsupplymaydiffer.Descriptively,youngchildrenwhoarenotinchild carearemuchmorelikelytobecaredforbyoneoftheparentsinthe home(oftenthemother),whileolderchildrenaremoreoftencaredfor byinformalchildminders,likerelatives,friends,ornannies.Thismay beduebothtoastrongerreluctanceamongparentsofyoungchildrento useinformalchildcarearrangements,ortolesssupplyofinformalcare foryoungchildren.Eitherway,wewouldexpectthattheavailabilityof childcareforyoungchildrenmayhaveastrongerpotentialtoincrease laborsupplyofparentsthanavailabilityofcareforolderchildren.
Ourresultsindicatethatchildcarefortoddlershassubstantialef- fectsonmothers’laborsupply.Amarriedorcohabitingmotherinduced touseafullyearofchildcarebythereformis32percentagepointsmore likelytobeemployed,comparedtomotherswhodonotusechildcare atall.2Thisisoverabaselineof63%participationbeforethereform, andassociatedwithanearningsincreaseof66,000NOK(8,000USD)for cohabitingmothers.Wealsoinvestigatepersistenceinthelaborsupply response,findingpositiveimpacts1–3yearslater.Forsinglemothers, wefindcontradictingandimpreciseresultsthatprecludestrongconclu- sions.
Proponentsofsubsidizedchildcarecommonlyclaimthatpartsofthe costofsuchsubsidiesareoffsetbytheincreasedtaxrevenuesorreduced benefitsgeneratedbytheadditionalincomeofworkingmothers.Using dataonactualtaxpayments,wecangobeyondback-of-the-envelope calculationsfoundinpreviousliteraturetoshowthattheincreasedtaxes causedbychildcareuseis,atleastinourcase,insignificantandrela- tivelysmall.Specifically,ourestimatessuggestanaveragetaxrateof about14%ontheextraincome,muchlowerthantheaveragetaxrate.
Atthesametime,theincreasedaccesstochildcareandhigheremploy- mentdoesnotseemtoleadtosignificantreductionsinbenefits,which wouldfurtherhavereducedpublicspending.3
Forfathers,wefind nolabor supplyresponse. This mayindicate thatmothersarestilltheprimarycaretakers,stayinghomewhenchild careisnotavailable.Meanwhile,estimatesonworkingagegrandpar- entsshowthatthereisnoresponseamongmaternalgrandparents,but modesteffectsonthelaborsupplyofpaternalgrandparents.Interest-
1 See,however,GivordandMarbot(2015);GouxandMaurin(2010)fortwo studiesonchildcarefor2-yearoldsandthestudiesontheQuebecreform(Baker etal.,2008;2015;Haecketal.,2015;KottelenbergandLehrer,2013;2017;
Lefebvreetal.,2009)discussedbelow,whichinvestigateschildcareforvarious agesfrom0to5yearsold.
2 Throughout,werefertomarriedandcohabitingmothersinterchangeably -ourmainsampleconsistsofbothmarriedmothersandunmarriedmothers livingtogetherwiththechildandthefatherofthechild,ascohabitationwithout marriageiscommoninNorway.
3 Notethatthisdoesnotincludethemechanicaleffectonthesubstantialcash- for-carebenefittiedtochildcareuseinsubsidizedchildcare,nordoesitinclude theparentalcopayment.Combined,theseimplyaparentalcostoffulltimechild careuseofaboutNOK90,000peryear,morethancancelingoutthepositive effectsonafter-taxincome.
ingly,whilepaternalgrandmothersworkmore,paternalgrandfathers worklesswhenthechildisusingchildcare.Thismaysuggestthatpa- ternalgrandmothersareimportantinformalcaregiverstosome2-year olds,whilethereisanindirectincomeeffectongrandfathers.
Wecontributetotherapidlygrowingliteratureestimatinghowchild careavailabilityaffectsparentallaborsupply.Mostofthepreviouslit- erature studiespre-school childrenatages 3–6,and findmodest ef- fects.4Foryoungerchildren,evidenceismorelimited,butsomerecent studiesindicatethateffectsmightbelargerforthisgroupthanolder children, inlinewith ourfindings.GouxandMaurin(2010), forin- stance,studyacutoff toschoolstartfor2-and3-yearoldsinFrance, estimating that about one singlemother entersemployment for ev- eryfour2-yearoldsenrolled.Incomparison,theydonotfindeffects on single mothersof 3-yearoldsor cohabitingmothers. Givord and Marbot (2015) study an increase in child care subsidies and esti- matethatthereformcausedsmallbutsignificantincreasesinthela- borsupplyof mothersof2-yearolds,butnot 3-yearolds.Carta and Rizzica(2018) studyareformthat expandedaccesstohighlysubsi- dizedchildcareto2-yearoldsinItalyinthemid-2000s,reportingef- fects of5–7 percentagepoints onthelabor participationofmothers.
BauernschusterandSchlotter(2015)studytheintroductionofalegal righttocarein Germanyfor3-yearoldsandfindimpactsof similar magnitudetowhatwefindinthecurrentpaper,whiletheestimatesfor aSpanishchildcarereformexpandingaccessfor3-yearoldsarearound halfaslarge(NollenbergerandRodríguez-Planas,2015).Estimatesfor theNetherlandsin Bettendorfetal.(2015)aremodest.Finally,esti- matesfor1–3yearoldchildrenarezeroornegligibleinstudiesfrom SouthKorea,GermanyandSweden(BusseandGathmann,2018;Lee, 2016;Lundinetal.,2008).
Animportantissue incomparing estimates acrossstudies, isthat theintensityoftreatment,andhencetheimplicitscalingofestimates, differswidely.GouxandMaurin(2010),forinstance,estimateafuzzy RD-designusingFrenchcensusdata,whereschoolenrollmentisinstru- mentedwithanagecutoff andtheoutcomeisstatedlaborforcepartic- ipation.Thisgeneratesestimatesperchildenrolledinschoolonaflow measureoflaborsupply.Incontrast,CartaandRizzica(2018)follow muchoftheliteratureonpreschoolagedchildrentoconsiderreduced formeffectsonyearlymeasuresoflaborsupplydirectly.Tointerpret theseestimates,theyconsiderchangesinchildcareaccessthatdonot capturetheintensityofuse.Sincenewchildcareplacesaccordingto theauthorswereopenedinSeptember,theutilizationratesinthefirst yearshouldbeexpectedtobemuchlowerthaninfollowingyears.In thiscase,theestimatesreportedwillunderstatetheeffectsofthechild careexpansiononlaborsupplyoverafullyear.
Our IV-strategyresolves theissue ofpartial treatmentandscales theeffecttotheincreaseduseofchildcarecausedbytheexpansion.
Thisshould,ontheonehand,ensurethatourestimatesarecloserto theexpectedimpactofthechildcareexpansionovertime,whenchild careplacesareavailablethroughouttheyear.Itshould,ontheother hand,facilitatetheapplicationofourestimatesinothercontexts,where thetake-upratesmaybeexpectedtodifferfromwhatweobservein our data. Our IV-estimatescan arguably be compared more directly tothoseinGouxandMaurin(2010),giventhatactualuseinFrance was substantial afterenrollment. Incontrast,estimates in Carta and Rizzica(2018)shouldbescaledbytheintensityofthereformtogive
4In the US, for instance economically small effects are found by Gelbach (2002), Cascio(2009), Fitzpatrick (2010); Fitzpatrick (2012)) and Barua(2014).SimilareffectsarefoundinseveralEuropeancountriesseee.g.
Finseraasetal.(2017);GouxandMaurin(2010);HavnesandMogstad(2011); Lundin et al.(2008), while severalpapers on a childcarereform in Que- becsuggestmoresizableeffects,seeBakeretal.(2008),LefebvreandMerri- gan(2008),Lefebvreetal.(2009)andHaecketal.(2015).Forreviewsofthis literature,wereferreaderstoBlauandCurrie(2006)andmorerecentoverviews inAkgunduzandPlantenga(2018)andMorrissey(2016).
ofchildcarearesmallerthanmightbeinitiallyexpected.Inthiscase, ratherthanreleasingmotherstothelabormarket,childcareistakenup bymotherswhoarealreadyworkingandrelyingonsomeformofinfor- malcarearrangement.Ourresultsthensuggeststhatinformalsourcesof caremaybelessimportantasanalternativetoformalcareforyounger comparedtoolderchildren.Bothalternativesimplythatthepotential forchildcarepoliciestostimulatemothers’laborsupplymaybelarger thansuggestedinmuchoftherecentliteratureonpreschoolchildren.
Thepaperproceedsasfollows:Wefirstcovertheinstitutionalset- tingandthechildcarereforminSection2,whileSection3presentsthe registrydatausedforestimation,thesamplesofinterestandsomede- scriptivestatistics.Section4presentsourIVmethodwithfixedeffects.
ResultsarefoundinSection5,includingpersistenceanalysis.Weper- formarangeofrobustnesscheckstosupportourestimatesinSection6, whileSection7concludes.Additionalresults,primarilyforfathersand othercaregivers,arefoundinanonlineappendix.
2. Institutionalsettingandthechildcarereform
AlthoughtherootsoftheNorwegianchildcaresystemdatebackto theearly19thcentury,5thesystemofuniversalchildcarewasintro- ducedafterWWIIasaresponsetoincreasingfemalelaborforce par- ticipationandthegoalofgenderequalityintheNordicwelfaremodel (MinistryofEducationandResearch,1998).Increasingexcessdemand forformalcareinthe60’sand70’sledtotheKindergartenActof1975, andastrongincreaseinthesupplyofformalchildcareforpreschool children(HavnesandMogstad,2011),eventuallyleadingtoahighcov- eragerateforpreschoolchildrenby1990.
Fig.1(a)showsthetrendsinchildcareaccessforpreschoolersand toddlersfrom1998–2012.Throughout,weuseasourmeasureofchild careaccessthereportedchildcarecoveragerate,i.e.theshareofchil- drenofaparticularagethatareenrolledinanychildcareonDecember 15thofeachyear.By2000,closeto80%of3–5yearoldswereenrolled informalchildcare.Atthesametime,childcareaccesswasmuchlower foryoungerchildren.In2000,lessthan50%of2-yearoldsand30%of 1-yearoldsinNorwaywereenrolledinchildcare,andtherewassub- stantialexcessdemand forchild care,seethediscussioninSection3 below.
Thisexcessdemandforformalchildcarewasthebackgroundforthe Kindergartenconcord,areformthatwasformallypassedintheNorwe- gianParliamentin2003withbroadbipartisansupport,butwhosemain lineswereagreeduponandmadepublictheyearbefore.Akeygoalof thereformwastofacilitateparentallaborforceparticipation,underthe premisethatuniversalchildcareiscentraltopromotinggenderequality inthelabormarket(MinistryofEducationandResearch,2002–2003) Thereformaimedtoofferaffordablechildcaretoallchildren,andto securequalityanddiversityinchildcareservices(MinistryofEducation
5 SeeMinistryofEducationandResearch(2008–2009)forathoroughtreat- ment.
lagof1–2years,astheywereappliedforanddisbursedaftertheslot wasopened.Thefigureshowsclearlyhowtotalinvestmentsincreased rapidlyfollowingthereform.Fig.1(c)showstheincreaseinstatesub- sidyratesperchildperyear.Fig.1(d)showsthechangesinthecompo- sitionofthecostscoveredbythemunicipality,thecentralgovernment andparentalfees.Itisclearthattheshareofcostscoveredbyparents hasdeclinedsignificantly.Thisfigurealsoshowsthatmunicipalsupport wasnotreducedasaresponsetotheincreasedgovernmentsubsidies.
Thelargeoverallincreaseinexpendituresovertheperiodisaresultof bothmorechildrenincare,highersubsidyrates,andanincreasingshare oftoddlers,requiringmorestaff andresourcesperchild.
A centralpart of thereform was theimplementation of a maxi- mum priceon childcare.This wasimplementedfrom 2004andput a capof 2,750NOKon themonthlyfeethatcouldbe chargedfrom parentsfor afulltimeslot.In2006,thecap wasloweredfurtherto 2250NOK permonth.7 Inaddition,all familieswith childrenbelow threeyearsoldwhowerenotenrolledinsubsidizedchildcarewereeligi- bleforasubstantialcashbenefitunderthecash-for-care(CFC)scheme.
Thisimpliesthatthepricetoparentswaseffectivelyabouttwice the levelsuggestedbythecap.8Throughouttheperiodweconsider,formal childcarewashighlyregulated.Tobeeligibleforthegeneroussubsi- dies,bothprivateandpublicchildcareinstitutionsweresubjecttostrict qualitycriteria,e.g.totheratioofpedagogicalstaff tochildren,open- inghours,parentalinvolvementandavailableplayingspaceperchild (KindergartenAct,2005).Themaximumpriceandthestrictregulation ofquality,ensuresthatformalchildcareinstitutionsarerelativelyho- mogenousintermsofobservableattributesofqualityandprice.
Theresponsibilityformeetingchildcaredemandfellonmunicipali- ties,whocouldexpandcarebydirectinvestmentorwiththehelpof private suppliers.Theexpansion,however,seems tobe hardtopre- dictforlocalgovernments,whooftenstruggledtomeetdemandover theyearsfollowingthereform.Themost commonreasonsforunder- supplyreportedbythemunicipalitiesthemselveswerea)demographic shocks,particularlyunexpectedchangesinthenumberofchildren,b) local geographicmismatch of supplyanddemand,c) unexpectedin- creasesindemandandd)unexpecteddelays inconstructionprojects (AsplanViak,2007).Manymunicipalitiesalsoseemtobeoverlyopti- misticinregardstocoveringallchildcaredemand,andwriteintheir annualreportsinseveralsuccessiveyearsthattheyexpecttoreachfull childcarecoveragethefollowingyear(AsplanViak,2004–2010).
6Allmonetaryvaluesaregiveninthousandsof2017-NOKunlessotherwise noted.
7Nominalvalues,comparablenumbersin2017NOKare3600NOKforthe 2004capand2800NOKforthe2006cap.NOK/USD≈8.4.
8OvertheperiodtheCFCbenefitvariedfrom2200to2800NOKpermonth in2017NOKperchild,equivalenttoapproximately270–350USDpermonth usingtheDecember2017exchangerate.Notethatthesechangeswerenation- wideandshouldnotaffectourestimationwhichisbasedonvariationacross municipalities.
AsillustratedinFig.1(a),thereformresulted inasharpincrease inmunicipalchildcareaccessfor1and2-yearolds.Overanineyear period,childcarecoverageratesfortoddlersincreasedbyaround40 percentagepointsfor bothagegroups.Inourempiricalanalysis, we aimtoevaluatetheimpactofthismassiveincreaseinchildcareaccess onthelaborsupplyofmothersandfathersusingthevariationintiming oftheexpansionbetweenmunicipalities.
3. Dataanddescriptivestatistics
Our datais basedon richadministrative registersavailable from StatisticsNorway,andcovertheentireresidentpopulation.Thedata containindividualinformationondemographics(e.g.sex,age,immi- grantstatus,maritalstatus,numberofchildren),socioeconomicstatus (e.g.yearsofeducation,income,taxespaid,employmentstatus),and municipalityofresidence.Incomeandemploymentdataarecollected fromtaxrecordsandotheradministrativeregisters.Thehouseholdin- formationisfromtheCentralPopulationRegister,whichisupdatedan- nuallybythelocalpopulationregistriesandverifiedbytheNorwegian TaxAuthority.Wealsohaveaccesstonationalregistrydataonmunic- ipalchildcareenrollmentreportedbythechildcareinstitutionsthem- selvesandaggregatedatthemunicipalitylevel.Importantly,thedata containuniquepersonalidentifiersthatallowustomatchchildrento parents,grandparentsandothercaregiversresidingwiththechild.Data onchildcareusecomesfromthecashforcare-registersasdetailedbe- low.WefinallyutilizevariousmunicipalitycharacteristicsfromStatis- ticsNorway,includingdataonruralandurbanpopulation,employment bysectorandgender,politicalrepresentationandmunicipalincomeand spending.
Todefinethepopulation ofinterest,we startwithalltwo-year oldsresidinginNorwayin2002–2008.Followingmostoftheliterature, wefocusattentionontheyoungestchildinthehousehold,andexclude multiplebirthsandchildrenwithyounger siblingsborntothesame motherorfather.9Wealsoexcludechildrenwithunknownmothersand ahandfulofhouseholdswithmorethanonechildinthesample(bornto differentparentsbutlivinginthesamehousehold).Thisleavesuswith asampleof325,396children.
Wenextidentifyallworking-agehouseholdmembersofthesechil- dren(ie.aged18to67).Weconsidercohabitingmothersandfathersto beparentsthatliveinthesamehouseholdasthechild.Noticethatour definitionofcohabitingparentsincludesbothmarriedandnon-married couples.Ifbothparentsarenotpresentin thechild’shousehold,we identifysuitablestepfathersorstepmothersfromage,familyrelations andgenderofthepresentparent.Thisallowsustoidentifythelikely caregiversrelevantforthevastmajorityofthechildreninthesample.
Frompopulationregisters,wecanalsoidentifymothersandfathersthat donot residewiththechildren,aswellasgrandmothers andgrand- fathers.Ourmainsamplesofinterestwillbecohabitingmothersand fathers,aswellassinglemothers andnon-residing fathers.Wehave alsoconsideredotherpotentialcaregivers,likestep-parents,butthese samplesaresmallanddonotgiveenoughpowertoallowmeaningful analysis.
Tomeasureindividualchildcareuse,weexploittheadministrative registersofcash-for-care(CFC)recipients,availablesince1999.Asdis- cussedabove,undertheCFC-scheme,allfamilieswithchildrenbelow threeyearsoldwhowerenotenrolledinsubsidizedchildcarewereeligible forasubstantialcashbenefit.FromtheCFC-registers,weknowwhich childrenreceivetheCFC-benefiteachmonthandtheexactamountdis- bursed.Aslongaseligibleparentstakeupthebenefit,wecaninferthe childcareuse ofeachchild.Becausechildrenwhoattendsubsidized careparttimeareeligibleforparttimecashforcare,wecanalsomea-
9 Fertilitycouldbeconsideredendogenoustothechildcareexpansion.In Section6,weincludealsochildrenwithyoungersiblingstoverifythatthis samplerestrictionisnotdrivingourresults.
suretheintensityofchildcareuseeachmonth.Thisapproachhasthe advantageofgivingacomprehensivemeasureofchildcareuseoverthe year,toallowcorrectlyscalingtheimpactofchildcareexpansionsthat arenotimmediatelytakenup.Ourfinalmeasuregivesforeachchild thefractionoffull-timeequivalentmonthsofchildcareuseduringthe year.10Notethatchildcareusebythisdefinitionwillrelateonlytothe use ofsubsidizedchild care.This isconsistentwithourdefinitionof childcarecoverageratesandinlinewiththedefinitionsusedinmost ofthepreviousliteraturewhereunsubsidizedchildcareisincludedin thecategoryof“informalcare”.
Tomeasurenumberofavailablechildcareslotsineachmunici- pality,weusethenumberofchildreninchildcarebyagefrommunic- ipalreports,measuredonDecember15theachyear,whenenrollment intheregularcalendaryeariscompleted.Ourmeasureofchildcare accessconvertsthesenumberstocoverageratesbytakingthenumber of slotsasafractionof thepopulationofthesameage asmeasured onJanuary1stinthefollowingyear,aroundtwoweeksafterthehead countchildcaremeasure.ThesedataarereadilyavailablefromStatis- ticsNorway.11Similarmeasuresareusedregularlyintheliterature,see e.g.(Dustmannetal.,2013;HavnesandMogstad,2011;Nollenberger andRodríguez-Planas,2015).
Onepotentialissuewiththesemeasuresisthatinclearingmarkets, childcareaccesswillbedeterminedjointlybysupplyanddemand.Ex- ploitingvariationinchildcareaccessmaythenpickupdemandshocks, which wouldraise thepotentialforreverse causalityandestimation bias.Anecdotalevidencesuggests,however,thattherewaslargeunder- supplyofchildcaretotoddlersinNorwayintheperiodweconsider.
Ifso,thevariationinchildcareaccessweexploitshouldbedrivenby variationinthesupplyofchildcare,ratherthanbyincreasedutilization ofexistingchildcareplaces.Inordertoverifytheanecdotalevidence, we wouldideallylike tohavedataonthenumberofapplicationsin eachmunicipality.Unfortunately,thisdataisnotavailable.Asanalter- native,wesecuredaccesstosurveydataonwaitinglistscollectedon behalfoftheMinistryofEducation.Thesurveycollectedinformation frommunicipalitiesonthenumberofchildren0–2yearsoldwhowere onwaitinglistsforchildcareineachmunicipalitybySeptember20th over2004–2009AsplanViak(2004–2010).Byaddingchildrenenrolled andchildrenonwaitinglists,wecanconstructameasureofthechild careapplicationrate.12
Fig.2 plotsthechild carecoverage rateandourmeasure of the child careapplication ratefor 0–2yearoldsoverourestimationpe- riod. Throughouttheperiod wecan measure,theapplicationrateis substantiallyhigherthanthecoveragerate,bybetween10and15per- centage points,indicatingexcessdemand.Note thatthesedata were collectedforthecombinedgroupof0–2yearoldsonly,sodonotspeak directly totheinstrumentwewillusebelow(childcarecoveragefor 2-yearolds).However,this islikelytocauseustounderestimatethe levelofrationing,sincethevastmajorityofchildrenwillneitherap- plynorenrollinchildcareinthefirstyearoflife,whentheirparents areon parentalleave.13 Theonlineappendixprovidesfiguresof the
10Wethusconsiderafullyearofhalftimecare(about20hperweek)tobe equivalenttoonehalfyearoffulltimecare.Wehaveexperimentedwithother measures,withoutthissubstantiallychangingtheresults.Inpractice,parttime careisrelativelyrare,asshowninFig.6below.
11http://www.ssb.no/statistikkbanken,table04683.Inafewcasesamunici- palitywillhaveacoveragerateslightlyabove1becausechildrenfromneigh- boringmunicipalitiesattendcare.Thesehavebeenadjustedto1.
12In a contemporary report for the Ministry of Education, ECON Anal- yse(2004)concludedthathiddendemandfromparentswhodonotapplywhen theywantcarefortheirchildren,wasnegligible.
13In2002,Norwegianparentswereentitledto42(52)weeksofparentalleave with100%(80%)wagecompensation,expandedto43(53)weeksin2005and 44(54)weeksin2006.Parentsarefurtherentitledtooneyeareachofunpaid leaveinimmediatecontinuationofregularparentalleave.
Fig.1. Childcareaccess,investmentsandfinancinginthe2000’s.Note:Childcarecoverageratesaredefinedasthenumberofchildrenincaretotheoverall populationofchildreninthatagegroup.Investmentsreferstototalgrossinvestmentinthechildcaresectorfrommunicipalandinstitutionalreports.Yearlystate subsidiesperchildarethenationalsubsidiespaidtomunicipalitiesperchildincare.Compositionoffinancingarefrommunicipalandinstitutionalreports.Sources: StatisticsNorwayandregjeringen.no.
Fig.2.Childcareapplicationandcoverageratesfor0–2yearolds.Note:Ap- plicationratesareconstructedbyaddingthenumberofchildrenincaretothe reportednumberofchildrenonwaitinglistsforcareanddividingbythenumber ofchildreninthemunicipality.Notethatthenumbersarenotdirectlycompa- rabletoFig.1(a),becausetheyinclude0and1-yearolds.Sources:Reportsfrom AsplanViak(2004–2010)andStatisticsNorway.
distributionofrationingacrossmunicipalitiesandyearsforwhichwe canmeasure,indicatingthatmoretnan55%ofthemunicipality-years wecanmeasure,coveringaround90%ofthechildreninoursample, weresubjecttoatleastsomedegreeofrationing,afairshareofthem tomoresevererationing.Inourrobustnessanalysisbelow,weusethis
datatoinvestigatethepotentialforreversecausalitybyrestrictingfocus tomoreseverelyrationedmunicipalities.
Tomeasurelaborsupplyweexploittwoalternativedatasources.
First,weuseyearlyearningsfromwagesandself-employmentcollected fromtaxrecords.Thismeasureincludesparentalleaveandsicknessab- sencebenefits.Asoutcomevariables,weusebothearningsdirectly,as wellasdummyvariablesforlabor marketparticipationbasedon the basicamountsintheNorwegianSocialInsuranceScheme(usedtode- finelabormarketstatus,determineeligibilityforunemploymentben- efitsaswellasdisabilityandoldagepension).Specifically,wefollow HavnesandMogstad(2011)andconstructdummyvariablesforemploy- mentandfull-timeequivalentstatusthatequaloneifearningsexceed 2and4basicamounts,respectively,andzerootherwise.In2017,one basicamountwasaround94,000NOK,orapproximately11,300USD.
Thetaxrecordsadditionallyprovidedataontotaltaxpayments,trans- fers andbenefits thatweuse asadditionaloutcomestoevaluatethe impactonpublicspendingandincome.
Second,weuse datafromthematchedemployer–employeeregis- ter,withinformationaboutallpublicemployeesandforabout80%of privateemployees.Thesedatagiveinformationonstartandenddates ofemploymentspells,andbracketedinformationoncontractedhours ineachspell.Fromthesedata,weconstructyearlymeasuresoflabor supplyasthenumberofweeksduringtheyearwhencontractedhours areabovefourhoursandabove30h.Acaveatinusingtheseoutcomes isthatwesetlaborsupplytozerowheninformationismissinginthe data.Thiswilltendtodrivedowntheleveloflaborsupplycomparedto thetruelevel.Thisshouldnot,however,affectourestimatesunlessthe patternofmissingobservationischangingovertimeandthesechanges arecorrelatedwithchangesinchildcarecoveragerates.
Table1
Descriptivestatisticsforallchildrenatage2,2002–2008.
Children
Variable All years All years SD 2002 2004 2006 2008
Female 0.49 0.50 0.49 0.49 0.49 0.49
Older maternal siblings 1.00 1.06 1.01 0.99 0.99 0.97
Child care use 0.47 0.42 0.32 0.39 0.51 0.67
Child care access 0.66 0.16 0.51 0.59 0.73 0.83
Observations 325,396 48,603 45,363 45,939 47,279
Table2
Descriptivestatisticsforcaregiversatchildage2.
Cohabiting Cohabiting Single Non-residing mothers fathers mothers fathers A. Outcome variables
Earnings 260.2 518.7 119.7 322.0
Tax 66.11 155.4 27.93 93.57
Transfers, excl. cash for care 56.74 18.49 198.1 50.51
Employed 0.678 0.914 0.309 0.699
Full-time eq. 0.398 0.841 0.150 0.539
Weeks of 4 h 37.17 43.58 21.35 32.17
employment above 30 h 26.44 41.69 13.15 29.02 B. Control variables
Age 32.65 35.55 29.52 32.46
Immigrant 0.097 0.074 0.087 0.080
Years of education 14.53 14.19 12.60 12.33
Number of children 2.029 2.056 1.719 1.818
below 6 years old 1.517 1.516 1.282 1.331
below 13 years old 1.893 1.878 1.553 1.596 below 18 years old 1.995 1.987 1.666 1.718
Observations 285,860 285,670 33,564 29,545
Note:OutcomeandcontrolvariablesaredefinedinSection3.Monetaryvaluesin1,000s of2017NOK.
3.1. Descriptivestatistics
Summarystatisticsforthesampleofchildrenandthefourmainsam- plesofcaregiversaregiveninTable1and2.Weseethatthesampleof childreniswellbalancedwithrespecttogender,andthatchildrenhave oneoldersiblingonaverage.Theaveragechildcareuseis0.47,cor- respondingto5.6monthsoffulltimecare.Table2givesdescriptive statisticsforfourgroupsofcaregivers:cohabitingmothers,cohabiting fathers,singlemothersandnon-residingfathers.Wenoticeimmediately inpanelAthesubstantialdifferencesintermsoflaborsupply.Single mothersareleastattachedtothelabormarket,withonly31%employed and15%infulltime-equivalentemployment.Incomparison,cohabiting fathersarestronglyattachedtothelabormarket,with91%employed and84%fulltime.Alsocohabitingmothersaremuch moreattached tothelabormarketthansinglemothers, withabout68%employed, and40%infulltime.Thesedifferencesaremirroredintheirearnings, wherecohabitingfathersonaveragemakealmosttwicethatofcohab- itingmothers,andmorethanfourtimesthatofsinglemothers.Wealso notethatlaborsupplymeasuredintermsofthenumberofweekswith contractedhoursabove4and30hoursreflectwelltheoverallpicture fromtheearnings-basedmeasures.
PanelBofTable2presentsmeansofcontrolvariablesforthefour groupsofcaregivers.Wenotethatcohabitingmothersandfathershave about14 years of completededucation. Thetable alsoshows single mothersandnon-residingfathershavesubstantiallylesseducationand arearoundthreeyearsyoungerthancohabitingparents.Meanwhile,fa- thersareaboutthreeyearsolderthanmothers,andsomewhatlesslikely tobeofimmigrantbackground.
Togetafirstlookatthetrendsinlaborforceparticipationamong thesegroups,Fig.3investigateslaborforceparticipationovertheperiod forthefourgroupsof caregiversseparately.Forcomparison,wealso includethetrendforworkingagewomenwithnoschoolagedchildren.
Wenotethatjustover60%ofthesewomenareemployed,of which
abouttwothirdsisfulltime.Forbothmeasures,however,theoverall trendisrelativelyflatovertheperiod.
Ifthelargeincreaseinchildcareavailabilityhasaneffectonparents’
laborsupply,wewouldexpectthatthiswasevidentinthefigurebyan increase inlabor supplyrelativetotheobservedpattern fortherest of thelaborforce.Forcohabitingfathersin oursample,thetrendis verysimilartotheoveralltrend,withlittlechangeovertime,though withemploymentabove90%,thelevelisclearlyhigher.Incontrast, cohabitingmothersexperiencedsubstantialincreasesinthelaborforce participationovertheseyears.Inparticular,employmentratesincreased byabout10percentagepoints,drivenlargelybyfulltime-equivalent employmentamongcohabitingmothers.Forsinglemothers,thereisalso aslightupwardtrend,butthepictureislessclear.
Finally,Table3presentsmeansofcontrolvariablesforcohabiting mothersbienniallyoverourestimationperiod2002–2008.Wenotethat mostvariablesexhibitarelativelyflattrendovertime,whichsuggests thatthecompositionofmothersisstableovertime.Theexceptionsare thegrowthinimmigrantbackgroundfromabout8%–12%andasmall increaseineducation.Similarpatternsareseenfortheothergroupsof caregivers,whicharereportedintheonlineappendix.
4. Empiricalstrategy
Themost straightforwardwaytoestimate theeffect ofchildcare onlabor supplyistoregressameasureoflaborsupplyonchildcare use.Thisignores,however,thatchildcareuseisendogenoustoparents’
laborsupplydecisions.Clearly,parentsthatenrolltheirchildreninchild carearelikelytobemorecloselytiedtothelabormarket.Thissimple approach istherefore likelytoyieldestimates ofhowchild careuse affectslaborsupplythatarebiasedupwards.
Imagineinsteadasocialexperimentthatrandomizedchildcareac- cessatthemunicipallevel.Thisrandomizationbreaksthecorrelation between child care access andunobserved determinants of parental
Fig.3.Laborforceparticipationforcaregiversof2-yearoldsandworkingagewomenwithoutchildrenbelow18,2001–2009.Note:Employmentandfulltime- equivalentisdefinedasearningsabove2and4basicamounts(BA),respectively,seeSection3.1BA≈NOK94,000≈USD11,300.
Table3
Descriptivestatisticsforcohabitingmothers,2002–2008.
2002 2004 2006 2008
A. Outcome variables
Earnings 221.3 240.7 271.2 313.7
(181.9) (230.0) (211.0) (220.1)
Tax 57.8 64.3 67.8 76.1
(63.8) (70.5) (74.7) (138.2)
Transfers, 66.8 58.8 52.9 47.7
excl. cash for care (57.9) (64.0) (60.2) (58.9)
Employed 0.63 0.65 0.70 0.75
(0.48) (0.48) (0.46) (0.43) Full-time eq. employment 0.35 0.37 0.42 0.47
(0.48) (0.48) (0.49) (0.50)
Weeks of 4 h 35.0 36.3 37.8 40.0
employment above (22.4) (22.2) (21.3) (19.9) 30 h 23.5 25.4 27.3 29.9
(24.6) (24.8) (24.7) (24.4) B. Control variables
Age 32.3 32.6 32.8 32.9
(4.81) (4.83) (4.83) (4.92)
Immigrant 0.084 0.090 0.10 0.12
(0.28) (0.29) (0.30) (0.32) Years of education 14.1 14.4 14.7 14.9
(2.93) (2.93) (2.95) (2.97) Number of children 2.05 2.03 2.03 2.00
(0.99) (1.00) (0.98) (0.98) below 6 years old 1.51 1.52 1.52 1.52
(0.58) (0.58) (0.58) (0.58) below 13 years old 1.91 1.89 1.89 1.87
(0.84) (0.83) (0.83) (0.82) below 18 years old 2.01 2.00 1.99 1.97
(0.94) (0.94) (0.92) (0.92)
Observations 42,587 39,648 40,444 41,817
Note:Thetablegivesbiennualmeansandstandarddeviations(inparentheses) intheestimationsample.VariablesaredefinedinSection3.Monetaryvalues in1,000sof2017NOK.
laborsupply.Comparinglaborsupplyofparentsinmunicipalitieswith andwithoutchildcareaccesswouldgiveareducedformestimateofthe effectofchildcareaccessonparentallaborsupply.Comparingchildcare useinmunicipalitieswithandwithoutchildcareaccesswouldgivea firststageestimateoftheeffectofchildcareaccessonchildcareuse.
Takingtheratiobetweenthetwo,wewouldgetanIV-estimateofthe effectofchildcareuseonparentallaborsupply.
TheintentionofourIV-approachistomimicthishypotheticalexper- iment.Weexploitthestaggeredexpansioninchildcarefollowingthe 2002childcareconcord,whichgeneratedlargespatialandtemporal variationinchildcareaccess.Thedistributionacrossmunicipalitiesof
childcareavailabilityfor2-yearoldsisillustratedinFig.4,wherewe drawmunicipalchildcarecoverageratesover2002–2008.Thefigure showsboththestrongincreaseinchildcareaccessovertheperiodwe consider,andthelargevariationacrossmunicipalities.
Westartbyconsideringthereducedformeffectofthisexpansion, relatingchangesinchildcareaccesstochangesinlaborsupply.This suggeststhefollowingmodel:
𝑦𝑖𝑘𝑡=𝜅𝑘+𝜂𝑡+𝛿𝐶𝐶𝑘𝑡+𝐗𝑖𝑘𝑡𝜸+𝜁𝑖𝑘𝑡 (1) whereyiktisameasureoflaborsupplyforcaregiveriinmunicipalityk inyeart,𝜅kand𝜂taremunicipalityandyearfixedeffectsandXiktisa vectorofparentandchildcontrols.Theparameterofinterest𝛿captures thereducedformeffectofincreasesinchildcareaccessonparents’la- borsupply.Givenexogeneity,i.e.thatchangesinaccessareasgoodas randomtochangesinlaborsupply,thisgivesanunbiasedestimateof theaverageimpactofanadditionalchildcareplaceonthelaborsupply of parents.Such reducedformestimationwillusuallybemost infor- mativeaboutthereturntothepublicinvestment,andisassuchoften regardedtogivethemostpolicyrelevantestimates.Indeed,thisisthe marginthatisestimatedinmostpapersintheliteratureonhowchild careaffectsparentallaborsupply,andwepresenttheseestimatesbelow forcomparisonwiththeexistingliterature.
Thereareatleasttworeasonswhyaccountingfortheintensityis important.inoursetting.First,newchildcareplacesaretypicallycre- atedinAugust,atthestartoftheschoolyear.Thismechanicallygives amaximumchildcareuseofaround42%formostchildren,i.e.5out of 12months.14This shouldleadthereducedformtounderestimate theimpactofafullyearofchildcareaccess.Second,totheextentthat childcareplacesarenotutilizedtocapacity,childcareinstitutionshave quiteextensivediscretiontoscaleprovisiontoactualuse,bye.g.adjust- ingemployeehours.Inthiscase,treatingchildcareplacesasuniform independentofutilizationwilloverestimatethecostofchildcareprovi- sion.Notethatthiswouldimplythatitisnotcorrectinourcasetouse thereducedformestimatestoevaluatecostefficiency.15
Giventhatnewchildcareplacesaffectlaborsupplyofparentsonly byaffecting theuseof childcare amongparents,wecan use anIV- approachtogetanestimateoftheimpactofchildcareonlaborsupply whichaccountsfortheintensityoftreatmentfollowingtheexpansion.
ThissuggeststhefollowingIV-strategy:Foreachmunicipalityandevery
14Thiscouldinprinciplebetakenintoaccountdirectlybyscalingtheestimates tothepotentialtake-up,i.e.dividingthereducedform-estimatesby5/12.
15Ofcourse,wecouldscalethecoststoreflecttheintensityratherthanthe estimates,butthiswouldinanycaserequireafirststageestimatefortheinten- sityofuseandwouldintroducestatisticaluncertaintyalsointhecostmeasures, complicatingtheanalysis.
Fig.4.Thedistributionofchildcareaccessfor2-year-oldsacross municipalitiesover2002–2008.Note:Thefiguredrawsthemean childcarecoverageratenationallyovertime(bulletsanddashed line)andthedistributionacrossmunicipalities(bars),weighted withpopulationsize.Data,estimationsampleandvariabledefini- tionsarediscussedinSection3.
year,weinstrumentindividualchildcareuseovertheyear,mikt,with thechildcarecoveragerate,CCkt.Specifically,weestimatethefollow- ing2SLS-modelinoursampleof caregiversof 2-yearoldchildren:16 𝑦𝑖𝑘𝑡=𝛼𝑘+𝜏𝑡+𝛽𝑚𝑖𝑘𝑡+𝐗𝑖𝑘𝑡𝜃+𝜖𝑖𝑘𝑡 (2)
𝑚𝑖𝑘𝑡=̃𝛼𝑘+̃𝜏𝑡+𝜋𝐶𝐶𝑘𝑡+𝐗𝑖𝑘𝑡̃𝜃+̃𝜖𝑖𝑘𝑡 (3) where𝛼kand𝜏taremunicipalityandyearfixedeffects.Notethatthe municipalityandyearfixedeffects,alongwiththerationingofthechild care marketdiscussed above,ensures that weexploit variation only fromnewlyopenedchildcareplaces.Inourbaselinemodel,weinclude controlsforthevectorXiktwhichcontainschildandcaregivercharac- teristics:Childcharacteristicsincludedummiesforgenderandmonth ofbirth,thenumberofoldersiblings,anddummiesforbothparents’
levelofcompletededucation.17Caregivercharacteristicsincludeage, agesquared,adummyforimmigrantbackgroundandthenumberof childrenbelowages6,11and18.Wealsoincludedummiesforlevelof educationtocontrolflexiblyforreturnstoeducation,includingpotential sheepskineffects.Descriptivestatisticsforthesevariablesarereported inTable2.Inourrobustnessanalysis,weshowthattheoverallpattern ofourestimatesdoesnotdependontheinclusionofcontrols.Standard errorsareclusteredatthemunicipalitylevelandrobusttoheteroskedas- ticity.
InthefirststageoftheIVmodel,weregressourmeasureofchild careuseovertheyearonmunicipalchildcareaccess.Ourmodelthere- foreaccountsfortheintensityofchildcareinducedbythechildcare expansionfollowingthereform.Fig.5(a)clarifiesthedatausedinour firststageinthemainsampleofcohabitingmothers.Intheleftpanel, weplotadummyforwhetherthechildusedanychildcareinDecember fromthecash-for-careregisterversusmunicipalchildcareaccessforour sampleofcohabitingmothers,whichisfrommunicipalreportsinthe samemonth.Datahavebeenbinnedbypercentile.Asisevidentfrom thefigure,thepointslineup closelyalongthediagonal,asexpected, andwithacorrelationcloseto1.18Thisindicatesthatthetwomeasures
16 AllmodelsareestimatedusingtheStatacommandreghdfe(Correia,2014).
17 Ratherthandroppingahandfulofobservationswithmissingeducationdata, weuseaseparatedummyforcaregiverswithmissingeducation.
18 Theslightvariationinthetwomeasurescouldbeduetoacombinationof measurementerror,movesduringtheyear,thatsomeeligibleparentsdonot applyforcash-for-care,thatoursampledoesnotincludechildrenwithyounger siblingsorthatsomechildrenattendcareinneighboringmunicipalities.
lineupverywellinDecember,verifyingthatourmeasureofchildcare useissound.
InthemiddlepanelofFig.5(a),weperformthesameanalysisre- placingthebinaryDecembermeasurewiththecontinuousmeasureof child careuseovertheentireyearthatweuse inourmainanalysis.
Thiscausesadownwardshiftwhichreflectsthattheaveragechildwho wasenrolledinchildcareinDecemberdidnotusechildcaretheentire year.Specifically,thechildcareplacescountedinDecemberwereuti- lizedonaverageonlyaround70%oftheyear.Thisaverageutilization ratewillreflectmostlychildrenwhoenrolledforthefirsttimeduring theyeartheyturn2yearsold,butalsothe(relativelyfew)childrenwho wereenrolledparttime.Thecorrelationisstillcloseto1,however,in- dicatingthattheaverageplaceisutilizedtoroughlythesameextentin municipalitiesbothwithlowandwithhighlevelsofchildcareaccess.
IntherightpanelofFig.5(a),wehavecontrolledformunicipality andyearfixedeffectstogetresidualizedmeasuresofchildcareuseand access.ThisreflectsthevariationthatweuseinourIV-model,coming fromnewlyopenedchildcareplaces.AsthesemostlyopeninAugust,it isnosurprisethatthechildrenwhooccupytheseslotsinDecemberuse lesschildcarethantheaveragetwo-yearoldincareinDecember,who mayhavestartedinthepreviousyearandattendedchildcarethrough thespring.Specifically,theassociationintherightpanelindicatesthat anewlyopenedchildcareslotcausesaboutonehalfyearofchildcare useinthatyear.19
To understand more aboutthe take-upof child care,Panel A of Fig.6showsthechildcareuseofchildreninoursamplebetweenages 14and35months,separatelyforchildrenbornindifferentmonths.For expositionalclarity,wehavehighlightedchildrenborninJanuary.We see thatthesechildrenarealmostuniformlymorelikelytousechild careastheygrowolder.Importantly,weseethatthereislittletake-up overtheyear,withthevastmajorityofchildrenstartingchildcarein August:ForchildrenborninJanuary,about20%startchildcareinAu- gustoftheyeartheyturntwoyearsold(whentheyare20monthsof age)andafurther20%inAugustoftheyeartheyturnthreeyearsold (whentheyare32monthsofage).Incomparison,onlyabout5%start childcareintheintermediatemonths.
Aconcerncouldbethatmanychildrenusechildcareonlyparttime, andperhapsexpandtheiruseduringtheyear.InPanelBofFig.6,we
19Notethattheabovediscussionshowsthatwearenotregressingindividual outcomesongroupmeans,asinapoorlydesignedpeereffectsanalysis.Fora detaileddiscussionofwhythisshouldbeabadidea,seeAngrist(2014).
Fig.5. Individualchildcareuse(top)andmarriedandcohabitingmothers’laborsupply(bottom)vs.childcareaccess.Note:Scatterplotsofindividualchildcare use(top)andoutcomes(bottom)againstchildcarecoverageratesfor2-yearolds,binnedbypercentile.Forvariableslabeled“residual”,wehaveremovedyearand municipalityfixedeffects.
Fig.6. Patternsofchildcareuseacrossageandcalendarmonths.Note:Panela)plotstheaverageshareofchildcareusersatchildage14to35monthsofageby birthmonth.ThegraphforchildrenborninJanuaryishighlightedforexpositionalease.Panelb)depictssharesofchildrenusingpart-time,fulltimeornochildcare bycalendarmonthduringtheyeartheyturn2,inferredfromcash-for-caredata.“None” indicateszerohours,“Parttime” indicates0–40h,and“Fulltime” indicates 40ormorehours.
showtheshareofchildreninoursamplewhoareenrolledinparttime, fulltimeandnochildcareineachcalendarmonthoftheyeartheyturn twoyearsold.Again,wenotethatthevastmajorityofchildrenenroll inAugust.Furthermore,weseethatchildrenareusuallyenrolledinfull timecareandthattheshareofparttimeuseisrelativelysmallandquite stableovertheyear.
Toillustratethereducedformofourmodel,Fig.5(b)showscatter plotsforourthreemainoutcomesagainstchildcareaccessafterremov- ingmunicipalityandyearfixedeffects.Wenotethatthereisastrong positiveassociationbetweenchildcareaccessandlaborsupply,andthat therelationshipiswellapproximatedbyalinearfunction,especially forourmeasuresofemploymentandfulltime-equivalentemployment.
Specifically,thecorrelationintheleftpanelsuggeststhatthechildcare expansionincreasedearningsofcohabitingmothersbyaround45,000 NOK.Scalingthis bytheassociatedfirststagefromabove,yieldsan IV-estimateofaround89,000NOK,closetotheIV-estimatesreported below.
Therearetwopotentialselectionissueswhenestimatingourparam- eterofinterest𝛽.Thefirstisselectionongains:Ifparentsinsomemu- nicipalitiesrespondmorestronglytochildcareaccessthanothers,and thisiscorrelatedwiththechildcareexpansion,thenourestimatesof howchildcareaffectsparentallaborsupplywilldifferfromtheaverage effectinthepopulation.Itwillbeaconsistentestimatoroftheeffect forthesubpopulationthatisaffected, butnot forthewholepopula- tion.IntheterminologyofImbensandAngrist(1994),ourestimatewill bealocalaveragetreatmenteffect(LATE).Inthepresenceofhetero- geneoustreatmenteffects,ourestimatesreflecttheaveragetreatment effectamongcompliers,roughlyparentsofchildrenthattook upthe newlyavailableslots.Wethinkthisisapolicyrelevantgroupwhether ornottheyarerepresentativeofthepopulationatlarge.
Second,wecouldworryaboutselectiononunobservables:Ifexpan- sionsinchild careaccessare,forinstance,positively correlatedwith otherdeterminantsoflaborsupply,thenourestimateswillbebiased upwards.Notefirstthatthefixedeffectswillcontrolforalltimeinvari- antdifferencesbetweenmunicipalities,andforallcommontimeshocks.
Ourconcernsshouldthereforebefocusedonchangesinpotentialcon- founderswithinmunicipalities.
Themostimmediatethreattoouridentificationisarguablythatchild caredemandisdrivingthevariationin childcareaccessthatweex- ploit,whichmightbiasourestimatesupwards.InSection3above,we discussedthisissueatsomelength,andconcludedthatthisisunlikely, sincedemandissubstantiallyhigher thansupplyover theperiodwe study.Toinvestigatethisissuefurther,wenowconsiderthepatternsof expansionsofcareacrossmunicipalities.
Toillustrateourempiricalapproachwerecenter thedata sothat year0istheyearwiththelargestincreaseinchildcareaccessforeach municipality.Wethengraphthechangesinchildcaresupplyovertime alongsidechangesinchildcareuseandkeyoutcomevariablesaround year0,afterremovingmunicipalityfixedeffects.Tofacilitateinterpre- tation,wehave notremovedyeareffects.These trendsaredrawnin Fig.7.
Westartbyconsidering thetimingof eventsin panels(a)–(d)of Fig.7.Ifourstrategyissound,thenwewouldexpectthatchangesin thelaborsupplyofcaregiversshouldfollowchangesinchildcareac- cessandnotviceversa.Thefigureshowstheassociationbetweenchild careaccess(dashedline),ontheonehand,andchildcareuseandlabor supply(solidlines),ontheother.Byconstruction,weseeasubstantial increaseinchildcareaccessinyear0,withajumpofaround15percent- agepointsinthecoveragerate.Thiscorrespondstoalittleunderhalfof theoverallincreaseinchildcareaccessovertheperiodweconsider,so willrepresentasubstantialfractionofthevariationthatweuseinour estimationsbelow.Becausewehavenotremovedyeareffects,itshould notbesurprisingthatthetrendinchildcareaccessispositivealsoin otheryears.
Inpanel(a)ofFig.7,weseethatchildcareusefollowscloselythe trendinchildcareaccess,withasubstantialjumpinchildcareusein
year0.Inpanels(b)–(d),wefurtherseethatlaborsupplyalsoincreases atthesametimeforbothearningsandourmeasuresofemployment andfulltime-equivalent.Incontrast,weseenoindicationofshocksin theoutcomessystematicallyprecedinglargeexpansionsinchildcareac- cess.Onthecontrary,thetrendsintheoutcomesareremarkablysmooth bothbeforeandafteryear0,lendingsupporttoourempiricalstrategy.
Inpanels(e)–(h)ofFig.7,wepresentsimilargraphsforkeycontrolvari- ables.Sincethesecanbecontrolledfor,theydonotposeadirectthreat toourestimates.Strongcorrelationbetweenchildcareexpansionand observablefactorswould,however,raiseconcernaboutpotentiallyun- correlatedtimeshocks.Unlikechildcareuseandthelaborsupplyout- comesthatweconsideredabove,thecontrolschangelittlearoundyear 0.Thissuggeststhatthesevariablesarenotdrivingtheassociationbe- tweenlaborsupplyandchildcare.Theexceptionisyearsofeducation, wherethereisaslightincreaseatyear0.Inpractice,wewillcontrol forallofthesevariablesinourestimation,toguardagainstpotential omittedvariablesbias.
Next,wecheckthebalanceofoursampleacrossmunicipalitiesthat expandchildcareaccessatdifferenttimes.Tothisend,werunare- gressionofchildcareaccessfortwoyearoldsonmunicipalityandyear fixedeffectsandasetofmunicipalitycharacteristicsmeasuredin2001, beforethereform,interactedwithyeardummies:
𝐶𝐶𝑘𝑡=𝜌𝑘+𝜎𝑡+𝝋𝒕𝑽𝒌,𝟐𝟎𝟎𝟏+𝜇𝑘𝑡
where Vk,2001 is avector of pre-reform municipalitycharacteristics.
Fig.8plotscoefficientsontheinteractionsbetweenyearandasetof keycharacteristics(seefootnote29fordetails),denoted𝝋tabove.Asys- tematicrelationshipovertimewouldsuggestthatmunicipalitiesofpar- ticulartypeshaveadifferentexpansionprofile.Estimatesarepresented in Fig.8andshowthatthereislittlesystematiccorrelationwithini- tialcharacteristicsofthemunicipality.Unsurprisingly,lowinitialchild careaccessfor2-yearoldshasastrongrelationshipwiththerolloutof accessforthisagegroup.20 Itisreassuring,however,thattherollout appearstobemostlyuncorrelatedwithothermunicipalitycharacteris- tics.Inparticular,thereisnoindicationofarelationshipbetweenthe rolloutofchildcareandtheinitialleveloffemalelaborforceparticipa- tion.Aparticularconcernmightbethatourestimatesmaybeinfluenced bychangesinlabordemand.Fig.3showsthattheoveralltrendinem- ployment,asmeasuredbyemploymentofworkingagewomenwithout school-agedchildren,wasroughlyflatovertheperiodwestudy(asimi- larpicturecanbedrawnformaleemployment).Whileoveralleconomic trendscannotimpactourfixedeffectsestimatesinanycase,wecould worrythatdifferentialeconomicchangesacrossmunicipalitiesmaybe correlatedwithchildcareexpansionandcausebiasinourestimates.
Fig.8showsthatinitialdifferencesinmunicipalitycharacteristics,in- cludingincomeandunemployment,doesnotseemtopredictinstrument rollout.Finally,ourestimatesarestabletotheinclusionorexclusionofa locallabormarketcontrol(municipalemploymentshareofworkingage menwithoutschool-agedchildren).Wethereforeconcludethatdiffer- entialgrowthinlabordemandisunlikelytocausebiasinourestimates.
Finally,evenwiththelargesetofcontrolvariablesthatweincludein ourmodel,onemaystillworryaboutchangesinunobservabledetermi- nantsoflaborsupply.Afterourmainresults,wethereforefurtherprobe thevalidityofourempiricalstrategybyassessingthestabilityoftheIV- estimatestoalternativespecifications,andbyperformingplacebotests.
Thesespecificationchecksbyandlargelendsupporttoourestimation strategy.
5. Results
Table4showsthereducedformestimatesfromourmodelamongthe fourmaingroupsofcaregivers.Focusingoncohabitingmothers,anad- ditionalslotinchildcareleadstoincreasesinearningsofaround31,000
20Notethatthisrelationshipbecomesmechanicalwhenmunicipalitiesap- proachfullcoverage.
Fig.7.Eventstudygraphsforcohabitingmothers:Changesintreatment(topleft)andoutcomesaroundthetimeofthelargestgrowthinchildcareaccessNote:The figuregraphsoutcomes,treatmentandcontrolsovertimeafterremovingmunicipalityfixedeffects.Dataarerecenteredsothatyear0istheyearwiththelargest growthinchildcarecoverageratesforeachmunicipalityafternettingoutyearlyshocks.Theleftaxes(solidlines)issizedtogofrom−0.4to0.4standarddeviations ofthevariableinthesampleofcohabitingmothers,butlabelsstillindicateabsolutevaluesofthevariable.Thescaleofthechildcarecoveragerate(rightaxis,dashed line)isinpercentagepoints.Data,estimationsampleandvariabledefinitionsarediscussedinSection3.Moredetailsinonlineappendix.
Fig.8.Whatpredictstimingofchildcareexpansion?Note:Thefigureplotsestimatesfromaregressionofthechildcarecoveragerateonmunicipalityandtime fixedeffectsandasetofpre-reformcharacteristicsinteractedwithyeardummies.Plottedarethecoefficientsontheinteractionofthecharacteristics(measuredin 2001)withtime,whichcanbeinterpretedasthedifferenceinexpansionofchildcareforamunicipalitywithoneunithigherlevelofthatcharacteristicin2001 inaparticularyeartotheaverageexpansionthatyear.Freeincomepercapitaandpopulationhavebeenstandardizedsothattheunitsareinstandarddeviations.
Standarderrorsareclusteredatthemunicipality.Seefootnote29foralistofcharacteristics.
Table4
Reducedformestimatesoftheeffectofanextrachildcareslotonannuallabor marketoutcomes.
Cohabiting Cohabiting Single Non-residing mothers fathers mothers fathers
Earnings 31.0 ∗∗∗ 0.33 − 10.9 − 2.19
(5.71) (17.9) (10.3) (23.4)
221.3 477.2 102.1 304.9
Taxes 4.20 − 9.51 − 14.2 ∗∗∗ 5.57
(2.66) (10.2) (4.86) (10.9)
57.8 139.9 28.1 82.9
Transfers − 0.30 0.35 20.7 ∗∗ 6.84
excl. cash for care (2.03) (2.28) (8.41) (7.18)
66.8 17.4 209.5 47.2
Employed 0.15 ∗∗∗ − 0.011 0.0097 0.046
(0.017) (0.010) (0.033) (0.040)
0.63 0.92 0.29 0.72
Full-time 0.081 ∗∗∗ − 0.0055 − 0.036 0.037 (0.017) (0.012) (0.023) (0.039)
0.35 0.85 0.13 0.56
Weeks of above 4h 4.91 ∗∗∗ − 0.55 1.74 − 0.48
employment (0.56) (0.66) (1.72) (1.96)
35.0 43.4 19.5 31.7
above 30h 5.16 ∗∗∗ − 0.65 − 1.90 0.19 (0.62) (0.73) (1.50) (2.02)
23.5 41.6 11.5 28.8
285,860 285,670 33,561 29,272
Note:Reducedformestimatesoftheoutcomesaslistedinrowheadersontheinstru- ment(childcareaccess),fixedeffectsandcontrols.Pre-reformmeansofoutcomes areunderlinedandrefertograndparentsof2-yearoldsin2002.∗(p<0.10),∗∗ (p<0.05),∗∗∗(p<0.01)
NOK, 15 percentagepoints increasedprobability of being employed and8percentagepoints increasedprobabilityof fulltime-equivalent employment.Asdiscussed,becausenewchildcareplacesareusually openedinAugust,theseestimateswillcaptureonlyapartyeareffectof theincreaseinchildcareaccess.Effectsonouryearlymeasuresoflabor supplyarethereforelikelytounderstatetheeffectofafullyearofchild careuse.Tocorrectlyinterprettheseestimates,weshouldcorrectfor theintensityofthetreatmentthatparentsexperience.
WethusturntoourIV-modeltogetestimatesthatmorecorrectly scalesthereducedformimpact.Baselineresultsfromequations(2)and (3)forthefourmainsamplesofinterestarereportedinTable5.Inpanel A,wereportestimatesfromourfirststage.Theinstrumentisrelevant andstrong,withF-statisticsontheexcludedinstrumentabove50inall samples.EstimatesinpanelAindicatethatnewchildcareplacesare utilizedonaverageabout43–47%ofafullyear,i.e.theequivalentof aboutfivetofiveandahalfmonthsinthatyear.Noticethatthescaling ofthereducedformestimateimpliedbythisfirststageestimateturns outtomatchwellwiththebasicscalingtopotentialtreatmentthatwe discussedabove.Thissuggeststhatutilizationofnewchildcareplaces isclosetosaturationinthemonthswhentheyareavailable.Looking acrossgroups,coefficients aresimilar,suggesting thatthetake-upof childcareslotsisrelativelyhomogeneousacrosschildrenwithdifferent livingarrangements.
InpanelBofTable5,wereportIV-estimatesforanumberofdifferent outcomes,whereestimatesonchildcareusearefromseparateregres- sionsforeachoutcome.Theestimatesshouldbeinterpretedastheeffect ofaddinganewchildcareplacewhichisutilizedtosaturationoverthe year.Ourfirststageestimatessuggestthatthedifferencebetweenthese estimatesandtheestimatesinTable4above,aremostlydrivenbythe additionofnewplacesinAugustratherthanatthestartoftheyear.Asa pointofreference,wealsoincludeinTable5themeanofthedependent variablein2002,priortothechildcareexpansion(underlined).21
21 OLS-estimatesarereportedintheonlineappendixforcomparison.These estimatesaremuchlargerthanourIV-estimates,inlinewithourintuitionthat motherswhoworkmorehavehigherdemandforchildcare.
OurIV-estimatessuggestsubstantiallaborsupplyresponsesamong the large group of cohabiting mothers. In particular, employment andfulltime-equivalentemploymentamongcohabitingmothersisesti- matedtoincreaseby32and17percentagepoints,respectively.Similar resultsarefoundforweeksofemploymentfromthematchedemployer–
employeeregister.Comparedtothepre-reformmeans,theseeffectsim- plyanincreaseofaround50%in themeanemploymentratesof co- habitingmothersoftwoyearolds.Thesestrongresponsessuggestthat childcareplaysanimportantroleingettingmothersofsmallchildren (back)intothelabormarket.Inlinewiththefindingsbelowofpersis- tenteffectsthefollowingyear,weinterprettheseresultsasdrivenby hasteningthereturntomoreorlessfulltimeworkratherthaneffects attheintensivemarginoflaborsupply.
Whenweconsiderearningsofcohabitingmothersdirectly,oures- timatesindicatethatchildcareusecausedanincreaseofabout66,000 NOKperyear,a30%increaseoverthepre-reformmean.22Atthesame time, we estimatethat taxespaid increase byaninsignificant 9,000 NOK,whilethereisnoeffectontransfers.Takingtheratioofthe(in- significant)estimateof9000NOKincreasedtaxestothe66,000NOK increasedearningsyieldsaverylowaveragetaxrateontheadditional earningsofaround13.5%,closetohalfoftheaveragetaxratesofthese mothersbeforethereform.Thisislikelycausedbynonlinearitiesinthe taxschedulethatmakesthemarginaltaxratedifferentfromtheaver- agetaxrate,particularlyforlowearnings.Thismeansthatsimpleback-
22Itiscommontotakelogsofthedependentvariableinordertointerpret theestimatesaspercentagechanges.Thepresenceofzerosintheoutcomeand thefactthattheshiftintoemploymentisanimportantmarginofresponseto childcaremakesthisunattractiveinourcase.Otheralternatives,relyingon variousmoreorlessarbitrarytransformationsoftheoutcomevariablethatin- volveaddingasmallnumbertoallobservationsbeforetakinglogs,inpractice putslargeweightsonsmallchangesclosetozero.Theinversehyperbolicsine transformationisarguablylessextremeinthiscase,butstillstronglyempha- sizeschangesinthelowerpartofthedistribution.Inlinewiththis,estimated elasticitiesforbothsingleandcohabitingmothersareverylargewhenweapply thistransformation.
roughlynull.Addtothistheco-paymentforchildcare,atabout40,000 NOKperyear,anditisclearthatchildcareisactuallyquitecostlyto parentsevenwhenincreasedearningsaretakenintoaccount.Thissug- geststhattherearenon-pecuniarybenefitsassociatedwithworkorthat futurebenefitsofreturningtoworkearlyaresubstantial.Whilewedo nothavedatatoinvestigatetheformer,weinvestigatepersistencein thelaborsupplyresponsebelow.
Turning to single mothers, the point estimates in column 3of Table5arenotsufficientlyprecisetoallowforstrongconclusionsabout thelaborsupplyeffects.Asweseebelow,however,themeanimpact seemstohidesubstantialheterogeneityovertheearningsdistribution, withpositive impactsat thebottom andnegativeeffectsat thetop.
Meanwhile,theestimatedeffectsontaxesandtransfersaresurprisingly large,withanestimateddropintaxespaidofabout30,000NOKandan increaseintransfersofnearly50,000NOK.Apotentialexplanationfor theseresultsisthatchildcareallowssinglemothersto(barely)meetthe activityrequirementslinkedtothetransitionalbenefitforsinglemoth- ers.23Note,however,thattheresultsforsinglemothersarelessrobust toourspecificationchecksbelow,suggestingthatweshouldbecareful inputtingtomuchemphasisontheseresults.
Estimatesfor non-residingfathersarenot sufficientlypreciseto drawfirm conclusions,butareingeneralclosetozero,lendinglittle supporttochildcareasanimportantdriveroflaborsupplydecisions forthisgroup.
Wehavealsoinvestigatedthelaborsupplyresponseofgrandpar- ents,underthehypothesisthatgrandparentsmayberelevantinformal caregiversforyoungchildren.Ifso,andifgrandparentsarestillattached tothelabormarket,wewouldexpecttoseeeffectsonthelaborsupply ofworkingagegrandparents.Estimatesforthesegroupdsarereported intheonlineappendix,andsuggestthatthereisnoresponseamongma- ternalgrandparents,butmodesteffectsonthelaborsupplyofpaternal grandparents.Interestingly,whilepaternalgrandmothersworkmore, paternalgrandfathersworklesswhenthechild isinchild care.This maysuggestthatpaternalgrandmothersareimportantinformalcare- giverstosome2-yearolds,whilethereisanindirectincomeeffecton grandfathers.
Wecanuseourresultstocalculatethenetcontemporaneousfiscal costtothegovernmentofanadditionalchildinfulltimecareatthetime ofenrollment.Wedothisforthecurrentyearonly,butnotethatwe estimatepersistenceinthelaborsupplyresponsebelow,findinglittle evidenceofincreasedtaxesinthefollowingyear,indicatingthatany impactofpersistenceonthesecalculationsshouldberelativelyminor.
Using2017-pricesand2008asthebaseyear,thestatesubsidiesforone fulltimechildincarewas107,500NOK.Inaddition,themunicipalities
23 Thetransitionalbenefit(“overgangsstønad”)ispaidtosinglemotherswho areatleast50%employed,inwork-relatedtrainingoractivelysearchingfora job.Thetransitionalbenefitis2.25timesthebasicamount,i.e.about210,000 NOKperyear.Thebenefitissubjecttoregularincometaxandisreducedby 45%ofanyincomeabove0.5basicamounts.
samebasicamountsweusedtoconstructourmeasuresofemployment (around94,000NOK).Specifically,weconstructsevenmutuallyexclu- sivedummyvariablesthatareequaltooneifearningsfallbelowone basicamount,betweenoneandtwobasicamounts,betweentwoand threebasicamounts,andsoon,withthelastcategorybeingearnings abovesixbasicamounts(around564,000NOK).Wethenestimateour IV-modelusingthesedummyvariablessuccessivelyasdependentvari- ables.Thisallowsustostudybinnedpartsofthemarginaldistribution ofearningstogetanimpressionofwhatpartsofthedistributionareaf- fectedbythechildcareexpansion.24Sincetheoverallimpactonfathers issmall,weconsideronlymothersinthisexercise.
Fig.9showsthemarginaldistributionin2002,i.e.themeanofour dummyvariables,asbullets.TheIV-estimateisrepresentedinbarswith associatedconfidenceintervals.Intheleftpanel,wereportestimates forcohabitingmothers.Thebullets(measuredontherightaxis)show thatpriortothechildcareexpansion,almost25%ofcohabitingmoth- ersearnedbelowonebasicamount,leavingthemessentiallyoutofthe laborforce.Fortheremainingcohabitingmothers,thedistributionof earningswasrelativelyflat,with10–15%ineachbracket.Incontrast, thedistributionforsinglemothersreportedintherightpanelofFig.9is heavilyskewed,with62%inthelowestearningsbracketandpopulation sharesbelow10%andsteadilydecliningaswemoveupthedistribution.
ThebarsinFig.9(measuredontheleftaxis)givetheestimatedim- pactofchildcareuseontheprobabilityofendingupineachbracket, andrevealstrongheterogeneity.Inparticular,cohabitingmothersseem tobeshiftingawayfrombothofthetwolowestbrackets,andintothe twomiddlebracketsclosetothethresholdforfulltime-equivalentem- ploymentthatweusedabove.Theinsignificantpointestimateonthe meanearningsofsinglemothersabove,turnsout tohidesubstantial heterogeneityovertheearningsdistribution.Estimates inFig.9sug- gestthatsinglemothersshiftawayfromthelowestbracketandinto thelowermiddlebrackets,pointingtowardsanincreasedincidenceof smallparttime(orlowwage)employment,butalsoawayfromthetop bracket,thoughestimatesarelesspreciseinthissmallersample.
5.2. Persistence
Thoughestimatedimpactsonearnings,transfersandtaxesdidnot supportthenotionofchildcareasapublicfinanceboonintheshort run,wemayquestionwhethertherearelongruneffectsthatcanhelp mitigatethecosts.Wethereforeinvestigatewhetherthereispersistence
24Notethattheevidencehereissuggestive,sincetheestimateswillreflect theeffectonthedistribution,notthedistributionofeffects.Thatis,though theestimatesaresufficienttoconcludeabouthowthechildcareexpansion affectedthedistributionoverall,andhenceformostwelfareanalyses,theydo notprovideinformationaboutthenumberofwinnersandlosersorthesizeof individualgainsorlosses,unlesswearewillingtomaketheboldassumptionof rankinvariance(i.e.noreshufflinginthedistributionwithtreatment).Seee.g.
KoenkerandHallock(2001)orKoenkeretal.(2017)forextensivediscussions oftheseissues.
