CHINHUI
JUHN
Universityof Houston
KEVIN
M. MURPHY
Universityof Chicago
ROBERT
H. TOPEL
Universityof Chicago
CurrentUnemployment,
Historically Contemplated
our BrookingsPaper"WhyHas the NaturalRate of
UnemploymentIncreasedover Time?" analyzed long-termchanges in
joblessness among Americanmen.' We documentedthe dramaticrise
between 1967 and 1989 in both unemploymentand nonparticipationin
the laborforce amongprime-agedmales. Ourmainconclusionwas thata
steep and sustaineddecline in the demandfor low-skilled workershad
reducedthe returnsto workfor this group,leadingto high ratesof unemployment,laborforce withdrawal,andlong spells of joblessnessfor lessskilled men. We found that time spent out of the labor force and time
spentunemployedaccountedin roughlyequalmeasurefor the long-term
growthin joblessness.We concludedthatstructuralfactors,primarilythe
declinein the demandfor low-skilledlabor,haddramaticallychangedthe
prospectsfor a returnto low ratesof joblessnessany time soon.
After thatpaperwas published,things appearedto change.The 1990s
openedwith a brief recessionthatwas followed by the longest sustained
decline in unemploymentin modernU.S. history. By the end of that
expansion,the unemploymentratehad reachedits lowest level since the
late 1960s, falling below 4 percentfor the first time since 1969. Some
macroeconomistsarguedthatthe so-callednaturalrateof unemployment
ELEVEN YEARS AGO,
We thank the Brookings Panel participants for helpful comments. Earlier versions of
this paper were presented at the National Bureau of Economic Research Labor Studies
meeting and at the University of Missouri. We acknowledge financial support from the
Center for the Study of the Economy and the State and the Milken Foundation.
1. Juhn, Murphy, and Topel (1991).
79
80
BrookingsPapers on EconomicActivity,1:2002
had permanentlyshiftedto 5 percentor below.2Because we had emphasized changesin the structureof labordemandthathad made a returnto
low ratesof joblessnessunlikely,these facts presenteda challengeto our
1991 framework.Maybe we were just wrong-maybe the demandand
supplyframeworkof ourpreviousworkis inconsistentwith ratesof joblessness in the post-1990 period. If so, we would join a distinguished
groupof social scientistswho have drawnattentionto a significantempirical phenomenononly to watchthatphenomenondisappearimmediately
As it turnsout, however,the frameworkthatwe developedfor
thereafter.3
thinkingaboutpre-1990 patternsof joblessness also does fairly well in
jobless time in the post-1990period.
helpingto understand
In this paper we look in some detail at employmentdata from the
1990s, revisitingissues raisedin ourearlierwork.Specifically,we ask:
-Have the trendswe identifiedin ourearlierpaper-the concentration
of nonemploymentamongthe less skilled,the growthof nonparticipation
in the labor force, and the increased durationof joblessness-been
reversedwith the fall in aggregateunemployment?
-Did the expansionof the 1990s reallyreturnthe U.S. labormarketto
conditionsof the late 1960s, as unemploymentstatisticsseem to indicate?
-Does the economicframeworkof supplyand demandwe utilizeda
decadeago still help in understandinglong-termdevelopmentsin unemployment,nonemployment,andlaborforce participation?
Ouranswersare surprising.First,the basic trendstowardlongerspells
of joblessness and rising nonemploymenthave continuedin spite of the
prolongedexpansionof nationaloutputandthe concomitantfall in unemployment rates. Long jobless spells and labor force withdrawalwere
more importantin the 1990s thanever before. Second, the fall in unemploymentto levels close to historicallows is very misleading.Broader
2. See, for example, Stiglitz (1997), Gordon (1998), and Staiger, Stock, and Watson
(2001).
3. Malthus founded the club. His theory that the forces of endogenous population
growth doomed the common people to perpetual poverty "explained" why incomes had
failed to increase over the period his data covered. Publication of Malthus's theory was followed by two centuries of almost continuous progress. More recently, when the returns to a
college education were at a record low in 1979, Richard Freeman (1976) offered a supplybased theory in The Overeducated American, only to see returns to a college education
increase steadily over the next fifteen years, reaching a record high. To Freeman's credit,
his model did predict a rebound, although not so large and sustained as the one that actually
occurred.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
81
measuresof joblessness show thatthe labormarketof the late 1990s was
morelike the relativelyslack labormarketof the late 1980s thanlike the
boominglabormarketof the late 1960s. Finally,the basic forces of supply and demand identified in our previous paper continue to have
explanatorypower.The theorydoes a reasonablygood job of explaining
those trendsthathave continued,as well as those thathave changed.
Recentdataalso provideconsiderableinsightinto what has happened
in the labormarketover the past decade.Overthe 1990s, even as unemploymentwas falling,time spentout of the laborforce was rising.In fact,
the increasein time spent out of the labor force was so large that total
joblessness-which combinesthe unemployedwith those who have withdrawnfrom the labor force-was as high at the business-cyclepeak in
2000 as it hadbeen at the previouscyclical peakof 1989, even thoughthe
unemploymentrate was roughly2 percentagepoints lower. In terms of
totaljoblessness, the often-praisedboom of the 1990s really represented
little in the way of employmentprogressfor Americanmales.
Althoughthe growthin the amountof time Americanmales spendout
of the labor force continuesa trendfound in our earlierresearch,other
featuresof the datachangedin the 1990s. The real wages of less-skilled
men, which had been falling steadilysince the early 1970s, stabilizedin
the 1990s andeven reboundedslightlyin the secondhalf of the decade.It
appearsthatthe thirty-yeartrendtowardgreaterwage inequalityhas run
its course,at least at the bottomof the wage distribution.The dataon joblessness reflect the impact of the changingwage trends.The long-term
divergencein employmentrates between low-wage workersand those
with higherwages, so pronouncedin our earlierwork, has stopped,and
unemploymentand wage gaps across skill groups have narrowed.The
congruencebetween patternsof change in wages and in employment
comportswith our previous work, which stresseddemand-drivenwage
changes as the dominantfactor driving secularchanges in employment
rates.
We are not the first to study the decline in unemploymentin the
1990s. Othershave emphasizedchangesin the compositionof the labor
force as a source of this decline. RobertShimerfound that aging of the
laborforce is importantin explainingthe decline in unemployment,particularly compared with the late 1970s.4 Lawrence Katz and Alan
4. Shimer (1998).
82
BrookingsPapers on EconomicActivity,1:2002
Kruegerinvestigatedto what extent the withdrawalof the incarcerated
populationfromthe laborforce, amongotherfactors,has led to a dropin
the aggregateunemploymentrate.'Otherpapershave exploredthe role of
improvementsin job searchtechnology.Forexample,DavidAutorargues
that temporaryhelp agencies may have helped improve the efficiency
with whichjob seekersarematchedwithemployers,thusbringingabouta
The arrivalof the Internetmay have
declinein frictionalunemployment.6
also reducedsearchcosts, althoughits impactis less certain.7
We show in this paperthata sharpdecline in the incidenceof jobless
spells accountsfor the lowerunemploymentratesof the 1990s,butthatat
the same time durationsof spells have remainedhigh. This fact is inconsistentwith a theorybuilt on decliningsearchcosts, which would imply
shorterunemploymentspells.
A relatedline of researchcomparesthe divergencein employmentoutcomes betweenthe UnitedStatesandEurope.Althoughboththe U.S. and
the EU economies may have experienced similar patterns of labor
demandduringthe 1970s and the 1980s, it is widely believed thatmoreflexiblelabormarketsandwages keptAmericanunemploymentratesrelatively low, while Europeanratesrose. Along these lines, severalpapers
emphasizethe importanceof interactionsbetweenmacroeconomicshocks
and labor marketinstitutions.8These papers find that althoughneither
macroeconomicvariables(oil prices, real interestrates,total factorproductivity,the laborshareof income)norlabormarketinstitutionvariables
(unemploymentbenefitsandduration,unioncoverage,collectivebargaining, employmentprotectionpolicies) alone can explain the differences
betweenthe UnitedStatesandEurope,a modelthatallows for interaction
effects fits the data well. But this shocks-plus-institutions
frameworkis
less successful in understandingrecentchanges in U.S. unemployment.
For example, Giuseppe Bertola, FrancineBlau, and Lawrence Kahn
5. Katz and Krueger (1999). They conclude that up to 0.4 percentage point of the rise in
the male employment-to-population ratio from 1985 to 1998 could be due to the bias from
ignoring the institutionalized population. For the sample of prime-aged males we study
here, the bias could be larger, further underscoring our finding that labor market conditions
did not improve much for prime-aged males in the 1990s.
6. Autor (2000a).
7. Autor (2000b); Kuhn and Skuterud (2000).
8. Bertola and Inchino (1995); Blanchard and Wolfers (2000); Ljungqvist and Sargent
(1998); Bertola, Blau, and Kahn (2001).
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
83
reportedthat the model significantlyunderpredictsthe decline in U.S.
unemploymentin the late 1990s.9
We revisit the evolution of joblessness in the United States, using
thirty-fouryears (1967-2000) of microdatafrom the CurrentPopulation
Surveys(CPS) conductedby the Bureauof the Censusandthe Bureauof
LaborStatistics.Ourmainconclusionsarethe following:
-Falling unemploymentrates over the 1990s greatlyexaggeratethe
improvementin labormarketconditionsfor prime-agedmales. Rates of
overalljoblessness-which includetime out of the laborforce-remained
roughlythe samein the late 1990s as they hadbeen in the late 1980s,even
as unemploymentrates fell. Rising labor force nonparticipationamong
prime-aged men largely offset declining unemployment,so that the
employment-to-population
ratioheld constant.
-Trends towardlongerdurationsof both unemploymentand nonemployment continuedin the 1990s, in spite of declining unemployment
rates.The probabilityof enteringunemployment(or nonemployment)fell
dramaticallyduringthe 1990s. The decline in the incidence of jobless
spells was so large that the likelihood of experiencingone reachedits
lowest level in the thirty-fouryearscoveredby ourdata.But therewas no
decline in the duration of unemploymentspells-these were about
2.8 weeks longerin 1999-2000 thanthey hadbeen a decadeearlier-and
the durationof nonemploymentspells increasedby over fourmonthsduring the 1990s. Broadlyspeaking,all of the long-termgrowthin joblessness is the productof longerdurationsof jobless spells.
-Although nonemploymentcontinuesto be concentratedamonglessskilled men, the trendtowardrisingjoblessness among the least skilled
reversedcourse somewhatin the 1990s. The largest declines in unemploymentoccurredamongmen in the lowest skill categories.Unemploymentamongmen in the bottom10 percentof the wage distributionfell by
4.6 percentage points between the cyclical peaks of 1989-90 and
1999-2000, while the declinein unemploymentat the medianof the wage
distributionwas about 1 percentagepoint. In contrast,over the longer
term the growth in nonemploymentis heavily weighted toward lessskilled men. Among men at the bottomof the wage distribution,the nonemploymentrate increasedby 13.5 percentagepoints between the late
9. Bertola, Blau, and Kahn (2001).
84
BrookingsPapers on EconomicActivity,1:2002
1960s and 2000, but by less than 1 percentagepoint for men with wages
abovethe medianof the distribution.
-The long-termdeclinein the real wages of less-skilledmen stopped
in the early 1990s and actuallyreverseditself slightlyin the latterpartof
the decade.Althoughthe wages of highly skilled men grew most rapidly
of all duringthe 1990s-continuing past patternsof relative growthinequalitybetweenmen at the bottomof the wage distributionandmen at
the mediancontractedslightlyover the decade.Overall,the trendtoward
greaterwage inequalityappearsto have stoppedfor males in the bottom
half of the wage distribution.
-Joblessness among less-skilled men has shown up increasinglyas
time spent out of the laborforce ratherthan as time spent unemployed.
Consistentwith our earlier work, we believe that this continuedtrend
towardlaborforce withdrawalreflectstwo factors:relativelylow returns
to work(realwages for the least skilledremainsubstantiallylowerthanin
the past) and increasinglyattractivenonworkopportunities,such as collecting disabilitypayments,which have shifted labor supply among the
least skilled.We findthatmorethan40 percentof the growthin nonparticipationis associated with an increase in men claiming to be ill or
disabled.
-Despite rising wages and rates of labor force participationfor
women,the highrateof joblessnessamongless-skilledmenis not the outcome of improvedlabor marketopportunitiesfor their workingwives.
Nonemploymentratesandratesof laborforce withdrawalincreasedmost
amongmen who did not have a workingwife. Looking acrossthe male
wage distribution,the proportionof men with a workingwife actuallyfell
amonglow-skilledmen, whose wages andemploymentrateswerefalling,
androse amongmen in the top 40 percentof the wage distribution,where
wages rose and employmentrates were stable. We conclude that longtermchangesin joblessnesshave been the resultof adverseshifts in labor
demand, perhaps coupled with policy-driven shifts in labor supply,
amonglow-skilledmen.
Data
Our data are drawnfrom the 1968-2001 AnnualDemographicFiles
that supplementthe MarchCPS. The CPS collects informationmonthly
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
85
from a rotating,randomsample of approximately50,000 U.S. households. It forms the basis for publishedgovernmentstatisticson earnings,
employment,unemployment,and laborforce participation,amongother
measures.Whereaspublishedlabor marketstatistics rely on questions
abouteach surveyrespondent'semploymentstatusin the referenceweek
of the survey(usuallythe thirdweek of the month),we studyretrospective information,collected each March,on labormarketoutcomesin the
previous calendaryear. Hence our data cover the thirty-fourcalendar
yearsfrom 1967 through2000.
In additionto personaland householdcharacteristicsfor each respondent, the retrospectivedata in the March survey recordthe numberof
weeks duringthe previousyear that the respondentworked,was unemployed, andwas out of the laborforce, as well as the respondent'snumber
of unemploymentspells. We measuretime spentunemployed(U) as the
percentageof the year spentin thatstate(for example,for the ith individual, Uiis the numberof weeks unemployeddividedby 52); time spentout
of the laborforce (0) andtime spentnonemployed(N = U + 0) aremeasuredin analogousfashion.This differsfromthe usualmethodof measuring time in unemployment,whichdividesweeks unemployedby weeks in
the labor force. Our method better summarizesthe allocation of time
across the three states, and it naturallyaggregatesacross individuals.'0
Using methodsdescribedbelow, we use informationon weeks worked,
unemployed,andout of the laborforce to calculateboththe incidenceand
the durationof jobless spells.
The survey also records a respondent'sannual earnings and usual
weekly hours workedfrom all jobs as well as occupation,industry,and
othercharacteristicsfor the longestjob held duringthe previousyear.We
use the informationon earnings,weeks worked,andhoursworkedto calculate averagehourlywages and to assign individualsa percentileposition in the overallwage distribution,as describedbelow. This allows us to
trackchangesin employmentoutcomes(U, 0, andN) for personsin differentpartsof the wage distribution.
We focus ouranalysison malesbecausethey were the focus of ourearlier work and because labor force participationissues for women are
significantlymore complex. To avoid issues associatedwith early retire10. Since the denominator for Ui, Ni, and Oi is always 52, the corresponding jobless
rate is simply the sample average of weeks in the state divided by 52.
86
BrookingsPapers on EconomicActivity,1:2002
ment, Social Security,and pensions,we focus on men who have one to
thirtyyears of potentiallabormarketexperience.For high school graduates this cutoffyields men who areroughlynineteento forty-nineyearsof
age, with correspondinglyhigher age intervals for those with more
schooling.We defineyears of labormarketexperienceas the smallerof
two numbers:age minusyears of educationminus seven, and age minus
seventeen."In addition,in orderto avoidmeasurementproblemsfor men
who spentpartof the year in school or in the military,we exclude those
who reportthat they did not work partof the year because of school or
militaryservice.
The employmentmeasureswe study are based on CPS respondents'
weeks worked, weeks unemployed,weeks out of the labor force, and
numberof unemploymentspells duringthe previousyear, as reportedin
the surveyweek. Using these data,we are able to identifythe fractionof
respondentswho experiencedsome unemploymentor time out of the
laborforce duringthe year, as well as the numberwho workedno weeks
duringthe year.We referto the latterevent as full-yearnonemployment.
Imputing Wagesfor Nonparticipants and Other Adjustments
We constructtwo samplesfor analysis. The "wage sample"contains
non-self-employedmen for whom valid observationsare available on
annualearnings,weeks worked,andusualweekly hours.'2Formen in the
wage sample,we calculatean hourlywage as the ratioof annualearnings
to the productof weeks workedand usual weekly hours.The "employment sample"includesthe entirewage sampleplus those men who lack
valid wage databecausethey did not work.For men not includedin the
wage sample,we imputea statisticaldistributionof wages basedon education,experience,andweeks worked.Foreach individualwith recorded
earnings,weeks, and hourswe projectthe log hourlywage on a quartic
function in potentialexperience,and we assign each individuala per11. We use age minus education minus seven rather the standard measure, age minus
education minus six, because age is measured at the survey week and we wish to measure
potential experience at the time of our wage and employment measures (which is the year
before the survey).
12. For the early years (before the 1976 survey) we impute usual weekly hours from
hours worked in the last week and individual characteristics, and we impute weeks worked
and unemployed from the categorical data based on averages calculated for the 1976-80
period.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
87
centile rankbased on his positionin the distributionof the residuals.For
personswith zero weeks workedin the previouscalendaryear,we impute
a wage distributionbasedon the observeddistributionof wages for those
who workedfrom one to thirteenweeks in that year.'3The imputation
assigns ten probabilityweights-each correspondingto the probability
that the individual'swage would come from a given decile of the wage
distribution-along with a meanwage for each decile.
Our 1991 papersoughtto explainchangesin jobless time by changes
in wages across skill categories. As we showed then, the relationship
betweencalculatedwages andtime workedduringa yearis contaminated
by measurementerrorin the latter,andthe relativeimportanceof this type
of measurementerrordeclines with the numberof weeks workedin the
previouscalendaryear.'4This builds in a negative relationshipbetween
labor supply (weeks worked)and calculatedwages, particularlyamong
men with high calculatedwages. As in our 1991 paper,we use data on
hourlywages for Marchrespondentswho were also in the outgoingrotation groupsto calculatethe wage adjustmentsthat would equatethe distributionsof calculatedretrospectivewages from the previous calendar
year and reportedhourly wages from the survey week. We then apply
these adjustmentsto each percentileof the wage distribution.The procedure effectively compresses the wage distributionin each year by an
amountthatwe attributeto measurementerrorin calculatedwages.
Armed with calculatedwages for those in the wage sample and an
imputedwage distributionfor those withoutvalid wage data, we group
individualsinto five "skill"categoriesbasedon theirpositionsin the wage
distribution.The percentileintervalsare 1-10, 11-20, 21-40, 41-60, and
61-100. As describedabove, each individual'swage percentileis calculatedbasedon his wages relativeto those of men with the samelevel of
experiencein a given year.Individualsin the wage sampleareassignedto
one of the five categoriesbased on theiractualwage, whereasthose with
13. Men with zero weeks worked resemble those with one to thirteen weeks worked in
terms of years of schooling completed and in terms of living arrangements (living alone,
with a spouse, or with other family). We also matched the outgoing rotation groups to the
March survey, yielding data on current (March) hourly wages for those who worked during
the survey week. Among individuals with zero weeks worked in the previous year but who
worked in the survey week, average log wages are nearly identical to those of men who
worked one to thirteen weeks in the previous year. See Juhn, Murphy, and Topel (1991) for
further details.
14. See Juhn (1992) for a more complete description.
BrookingsPapers on EconomicActivity,1:2002
88
imputed wages are assigned probabilities of being in each of the
categories.
Changes in Wages, 1967-2000
In light of the prominencewe assigned to wage changes in our 1991
paper,it is worthwhileto review what has happenedto both wage levels
and the distributionof wages since then. Figure 1 and table 1 summarize
the mainfeaturesof the data.Figure 1 shows trendsin real hourlywages
by wage percentilegroup(skill category)since 1967;the dataareindexed
to equal 100 in 1970. Forourpurposesthe most interestingaspectof these
datais thatwage inequalitystoppedincreasingin the 1990s, especiallyat
the lower end of the wage distribution,andthatthe real wages of all skill
categories increased after 1993. For less-skilled workers, real wage
growthin the 1990s representeda slight reversalof a twenty-yeardecline
in the returnsto work,whichhadfallenby nearly30 log pointsafter1972.
Figure 1. Real Wages, by Wage Percentile Group, 1967-2000a
Index,1970= 100
115 110
to 1
1050 -61
100
-
'
-
''''"
95
90
85
I to 40
80
75
II to20
to
I
1970
I
I
1975
1980
~ ~
~~~~~~~~I
1985
1990
1995
Source: Authors' calculationsbased on annualMarchCurrentPopulationSurvey (CPS) data.
a. Reportedhourly wage (in naturallogarithms)is projectedon a quarticfunction in potentialexperience. Men are assigned a
percentile category based on their position in the residual distribution.Wages for nonworkersand self-employed workers are
imputed.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
89
Table 1. Changes in Real Wages, by Wage Percentile Group, 1967-2000a
Percent
Wagepercentilegroup
Period
Ito10
11to20
21 to 40
41 to 60
61 to 100
1967-69 to 1988-89
1988-89 to 1994-95
1994-95 to 1999-2000
-24.8
-3.9
6.1
-19.9
-7.1
6.7
-12.7
-8.3
6.8
-4.0
-7.6
6.4
5.5
-3.6
10.7
1988-89 to 1999-2000
1967-69 to 1999-2000
2.2
-18.7
-0.4
-13.2
-1.5
-5.9
-1.2
2.4
7.1
16.2
Source: Authors' calculations using annualMarchCurrentPopulationSurvey (CPS) data.
a. Reportedhourly wage (in naturallogarithms)is projectedon a quarticfunction in potentialexperience. Men are assigned a
percentile category based on their position in the residual distribution.Wages for nonworkersand self-employed workers are
imputed.
Even so, average wages of the least-skilled men were roughly 20 percent
lower in 2000 than in the late 1960s (table 1), whereas those of men in the
top 40 percent of the distribution increased by a roughly equivalent
amount. As we showed in our 1991 Brookings Paper, wage declines were
most prominent among those whose employment outcomes are most sensitive to wage changes-the least skilled-whereas rising wages are concentrated among those with less elastic labor supply.
With this evidence as background, we turn to evidence on changes in
joblessness, both in the aggregate and across the wage percentile groups
defined above. We return to the implications of wage changes in the concluding section.
Unemployment, Nonparticipation, and Nonemployment
We begin by describing trends in unemployment and comparing our
March CPS-based data with unemployment statistics from the monthly
CPS published by the Bureau of Labor Statistics. Figure 2 shows that by
2000 the unemployment rate had reached its lowest level in thirty years,
and unemployment rates in 1999-2000 were close to the extremely low
rates seen during the late 1960s.15This is the culmination of a long downward trend in unemployment: in both the 1991-92 and the 2001-02 recessions (not shown), the peak unemployment rate was lower than the peak
in the preceding recession, reversing a trend of rising peaks across the
15. The publishedratehas been adjusteddownwardby 0.86 percentagepointto equate
the meansof the monthlyandthe Marchseriesover the sample.
BrookingsPapers on EconomicActivity,1:2002
90
Figure 2. AlternativeMeasuresof the UnemploymentRate, 1967-2000
Percent
10
Monthlysurveyb
Marc surve'
9
8
1111
7
6
5I-
March ~ surveyvr
h~smpe Mrc
4/
3
1970
1975
1980
1985
1990
1995
Source: Authors'calculationsbased on Bureauof LaborStatisticsand MarchCPS data.
a. Subsetincludingonly prime-agemales; calculatedas the numberof weeks unemployeddivided by 52.
b. Each observationfrom the monthly survey is reducedby 0.86 percentagepoint to equate the means of the monthly and the
Marchsurveys over the sample.
It appearsfrom the figure
1970-71, 1974-75, and 1982-83 recessions.m16
thatthe U.S. economyhas come full circle:unemploymentrose for fifteen
years (from 1968 to 1983) and then fell over the next seventeenyears
(from 1983 to 2000), with interveningcyclical swings. One might conclude from these data that the labormarketconditionsof the late 1960s
andlate 1990s were comparable.
Figure 2 also shows annualunemploymentrates for our sample of
prime-agedmales (calculatedfrom the MarchCPS dataas weeks unemployed divided by weeks in the labor force). Although the two series
shouldnot be identicalbecauseof differencesin the underlyingpopulations (oursampleconsists only of prime-agedmales, whereasthe overall
unemploymentdata are from the full populationof labor force participants aged sixteen and over), the two series are remarkablysimilarin
terms of underlyingtrendsand rankingsof cyclical variationsin unem16. The recession of 1980 did not fit this pattern, but as the figure shows, it did not represent much of a peak in terms of unemployment rates.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
91
ployment.One would reachthe same basic conclusionsaboutunemployment trendsfrom the publishedmonthlyseries as from our calculations
based on the MarchCPS data.Fromhere forwardwe analyzethe March
CPS dataexclusively.
A majorfinding of our 1991 paperwas that the long-termgrowthin
unemploymentgreatlyunderstatedthe growthin joblessness.Recentdata
suggest thatchangesin unemploymentshown in the aggregateand in the
CPS statisticsare even more misleadingfor the 1990s. This is illustrated
in figure 3, which plots two series: the fractionof annualweeks spent
unemployedandthe fractionof annualweeks spentout of the laborforce.
The nonemploymentrate is the sum of these fractions,so that the combined height of the two shadedregions representsthe proportionof the
year spent out of work. Figure 3 confirmsthat measuredunemployment
fell duringthe 1990s to levels comparableto those in the 1960s, but the
conclusion in terms of overall jobless time is much different:the late
1990s were much like the 1980s, in that the decline in unemployment
Figure 3. Unemployment,Nonparticipation,and Nonemployment,1967-2000
Percentof calendaryeara
16 -
Nonemploymentb
1412 10 -
4
1970
1975
1980
Source: Authors' calculationsbased on MarchCPS data.
a. Numberof weeks a year in indicatedstate divided by 52.
b. Unemploymentplus nonparticipation.
1985
1990
1995
92
BrookingsPapers on EconomicActivity,1:2002
over the 1990s is not reflectedin a lower overallrateof joblessness.This
means that, on net, men who left unemploymentdid not find jobs but
ratherleft the laborforce, so thatthe employment-to-population
ratiowas
unchangedfromits level in the 1980s.
Table 2 summarizesthe data in figure 3 by aggregatingthe data into
nine time intervalscorrespondingroughly to peaks and troughsin the
business cycle, as measuredby aggregateunemploymentrates. Unemploymentshows a strongcyclical patternas well as a long-runupward
trend(whethermeasuredpeakto peakor troughto trough)untilthe recession of 1982-83. After 1982-83 unemploymentratesfall or stay constant
(again whethermeasuredpeak to peak or troughto trough).In contrast,
the fractionof the year spent out of the laborforce rises betweenevery
pairof intervals.In fact, whereasthe unemploymentratein 1999-2000 is
very close to its level in 1967-69, the nonemploymentrateis 4.7 percentage pointshigher,andthe fractionof the year spentout of the laborforce
is roughlydoublewhatit was in 1967-69. It is difficultto concludefrom
these datathatemploymentconditionsof the late 1960s andthe late 1990s
were "similar"in any meaningfulsense.
Consider next the eleven-year interval between the business-cycle
peaks of 1988-89 and 1999-2000. Over this peak-to-peaktime spanthe
unemploymentratefell by 1.3 percentagepoints-from 4.3 percentto 3.0
percent-but the percentageof men who were out of the laborforce rose
by exactly the same amount,from 6.7 percentto 8 percent.This left the
nonemploymentrateat the same level (11.0 percent)in 1999-2000 as in
1988-89, even thoughthis period spans the longest sustainedeconomic
expansion,andthe largestdeclinein unemployment,on record.
We next divide the growthin nonemploymentalong a second dimension. The percentageof weeks spent out of work is equal to the sum of
two components:the fractionof men who did not workat all overthe year
(for whomthe fractionof weeks spentout of workis 100 percent)andthe
fractionof weeks spentout of workfor those who workedsome positive
amount(multipliedby the fractionof menwho workedat leastone week).
In what follows we refer to these two componentsas "full-yearnonemployment"and "part-yearnonemployment,"
respectively.This decompositionallows us to examinehow muchof the growthin nonemploymentis
accountedfor by men with very long stretchesof joblessness-that is,
spells thatareso long thatmendo not workat all duringa calendaryearandhow muchis due to men with "transitory"
jobless spells.
b. a.
Source:
Period
1972-73
1988-89
1982-83
1991-92
1978-79
1975-76
1971-72
1967-69
Number
1999-2000
of
Unemployed
Authors'
weeks
plus
in
Units
Table
as 2.
indicated
calculations
indicated
Peak Peak Peak Peak Peakbusiness
Phase
based Trough
Trough
Trough
Trough
nonparticipating;
of
status
on
Unemployment,
cycle
detailsMarch
divided
may
by
CPS
not52.
data.
sum
3.06.34.39.04.36.93.84.52.2
to
Nonparticipation,
Labor
Unemployed
totals
and
force
because
of
status
rounding.
8.07.56.76.35.95.65.04.94.1
(percent Nonemployment
of
Nonparticipating
during
calendar
13.811.0
15.2
11.0
10.212.4
8.89.46.3
yeara)
Nonemployedb
Business-Cycle
Peaks
and
-3.32.0-4.74.6-2.53.0-0.72.3
(percentage
Change
in
unemployment
Troughs,
points)
1967-2000
0.50.80.50.40.30.60.10.8
(percentage
Change
in
nonparticipation
points)
94
BrookingsPapers on EconomicActivity,1:2002
The results,shownin table 3 and figure4, are striking.The amountof
joblessness accountedfor by those workingat least partof the year was
only slightly higherin 1999-2000 than in 1967-69 (4.9 versus 4.5 percent). But the amountof joblessness accountedfor by those who did not
workat all over the yearmorethantripled,from 1.8 percentin the 1960s
to 6.1 percentin 1999-2000. Moreover,whereaspart-yearnonemployment declined by 4.5 percentagepoints from its recessionarypeak in
1982-83 to 1999-2000, full-yearnonemploymentincreasedslightly.This
is particularlystrikinggiven that the interveningperiodis characterized
by two of the longesteconomicexpansionson record.
Whatexplainsthis trendtowardlong-termjoblessness?Onepossibility
is that those men with the least favorablelabor marketprospectshave
simplydroppedout of the labormarket:the so-calleddiscouragedworker
effect. Figure5 addressesthis possibilityby disaggregatingnonemployment and nonparticipation,
respectively,by reportedreason.We distinthree
main
guish among
groups:those who reportedthat they could not
findwork,those who reportedthatthey wereill or disabled,anda residual
categorywe label "other."Overthe periodcoveredby ourdata,the figure
shows only a small increasein the proportionof men who reportedthat
theycouldnot findwork.Risingsharesof the "ill or disabled"and"other"
categoriesaccountfor the largest changes of both nonemploymentand
The largerimpactof the "ill or disabled"categoryis on
nonparticipation.
Table 3. Part-Yearand Full-Year Nonemployment,1967-2000a
Percentof calendaryear
Period
1967-69
1971-72
1972-73
1975-76
1978-79
1982-83
1988-89
1991-92
1999-2000
Phase of
business cycle
Part-yearb
Full-yearc
Total
Peak
Trough
Peak
Trough
Peak
Trough
Peak
Trough
Peak
4.5
6.5
6.0
8.4
6.8
9.4
6.5
7.7
4.9
1.8
2.9
2.8
4.1
3.8
5.8
4.6
6.0
6.1
6.3
9.4
8.8
12.4
10.2
15.2
11.0
13.8
11.0
Source: Authors' calculationsbased on MarchCPS data.
a. Details may not sum to totals because of rounding.
b. Fractionof males nonemployed for part of the year multiplied by the average percent of the calendaryear spent nonemployed for this group.
c. Fraction of males nonemployed for the entire year multiplied by the percent of the calendar year spent nonemployed
(100 percent).
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
95
Figure 4. Part-Yearand Full-YearNonemployment,1967-2000
Percentof calendaryear
16 j2i~~~~~~~~~Total
141412 10
8-
6
2
1970
1975
1980
1985
1990
1995
Source: Authors' calculationsbased on MarchCPS data.
a. Fractionof males nonemployedfor partof the year multipliedby the averagepercentof the calendaryear spent nonemployed
for this group.
b. Fractionof males nonemployedfor the entire year multiplied by the percent of the calendaryear spent nonemployed (100
percent).
the nonparticipation rate (middle panel of figure 5): men in this category
account for about 42 percent (0.8 out of 1.9 percentage points) of the
increase in nonparticipation between 1982-83 and 1999-2000. The rest is
"other."The bottom panel of figure 5 narrows the focus to men who were
full-year nonworkers, for whom the effect of rising disability is more
prominentstill. Virtuallynone of the long-termincreasein full-yearjoblessness is accounted for by discouraged workers: the "other" category
and persons reportingjoblessness for health reasons account for the secu-
lar increase.
A largeliteratureexaminesthe impactof changesin the disabilitybenefits programon the labormarketparticipationof male workers.17These
papersdocumenta substantialgrowthin the disabilityrolls in the early
17. Parsons(1980); Bound (1989); Bound and Waidmann(1992); Autorand Duggan
(forthcoming).
Figure 5. Nonemploymentand Nonparticipation,by ReportedReason, 1967-200a
Percentof calendaryearNonemployment
16 14 12 -
10
8
Ill or disabled
6
Nonparticipatiaoknvaiabl
9
87
6
5
4
Ill or disabled
3
2
91
Full-year nonemployment
9
8
7
65
43
Ill or disabled
2
1970
1975
18
Source: Authors'calculationsbased on MarchCPSdata.
a. Excludes individualswho reportbeing studentsor retired.
195
1990
1995
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
97
1970s, linked to the sharpdecline in participationamong older males.
Becauserealwages were risingfor the most partover this period,the earlierepisodeis consistentwitha reductionin laborsupplyin responseto the
improvingnonmarketalternativerepresentedby disabilitypayments.Durchangestighting the early 1980s,however,legislativeandadministrative
led
to
in new
these
standards
reductions
tighter
standards;
ened eligibility
awardsandterminatedbenefitsfor a substantialfractionof beneficiaries.
After 1984, eligibility criteriawere substantiallyliberalized,and this
led to increasedreceipt of disability payments.18 Examiningaggregate
time series as well as cross-statevariation,John Bound and Timothy
Waidmannconcludedthatvirtuallyall of the increasein nonemployment
amongthose reportingthatthey were ill or disabledin the CPS could be
explainedby increasedreceipt of disabilitybenefits.19Autor and Mark
Dugganconcludedthatliberalizationof eligibilityfor disabilityinsurance
interactswith adverse shifts in labor demand,as otherwiseemployable
over unemploymentor low-wage
men opt for subsidizednonparticipation
work.Figure6 offers indirectsupportiveevidenceon this point,comparing the changes in unemploymentand nonparticipationbetween peaks
and troughsof differentbusinesscycles. The figureshows thatincreased
nonparticipationaccountedfor a much largerfractionof rising nonemploymentin 1989-92 thanin earlierrecessions.The smallestcontribution
was in the recession of 1992, when eligibility rules
of nonparticipation
were tightened.The increasein nonemploymentamongthe ill or disabled
accountedfor nearly 16 percentof the total change in nonemployment
between 1989 and 1992 (not shown), a much higherproportionthan in
previousrecessions.Nonemploymentof the ill or disabledactuallyfell
during the 1982 recession, an observationthat also likely reflects the
tighteningof eligibilityrulesduringthis period.
Table 4 decomposes secular changes in nonemploymentbetween
1967-69 and 1999-2000, as well as over the 1990s. In the 1990s the data
indicate that roughlyhalf (0.8 percentagepoint) of the 1.5-percentagereflectsa shiftin laborsupplycausedby
pointincreasein nonparticipation
improving nonmarketalternativesto working. There is no reason to
believe thatthe healthof Americanmen deterioratedover the decade(and
Yet nonparticipation
caused
much reasonto believe that it improved).20
18. AutorandDuggan(forthcoming);BoundandWaidmann(2000).
19. BoundandWaidmann(2000).
20. See MurphyandTopel (2001), for example.
98
BrookingsPapers on EconomicActivity,1:2002
Figure 6. Changesin Unemploymentand NonparticipationEntering Recessions
Percentagepoints
5 -
0 Unemployment
E Nonparticipation
43-
2-
fAf?:$d9;
'S'
i;
1969-71
1973-75
.....
........
;.~~~~~~~~~~~~........
1979-82
1989-92
Source: Authors'calculationsbased on MarchCPS data.
by self-reportedhealthreasonsincreasedby 0.8 percentagepointover the
decade.Unlike in the early 1970s, when real wages were risingrapidlyas
nonparticipationincreased,real wages remainedlow and were falling
over the firsthalf of the 1990s. This fact makesit moredifficultto parcel
out the componentdue to shiftinglabor supply.Yet with the increasein
real wages in the latter half of the decade, continuinggrowth of nonparticipationindicatesa shift in labor supply. In a manneranalogousto
interpretationsof the Europeanunemploymentexperience,the dataindicate thatthe interactionof disabilitybenefitsandlabormarketshocksmay
be of key importance in understandingrising rates of labor force
withdrawal.2'
Figure7 summarizesour previousresults,showinglong-termchanges
in three alternativemeasures of joblessness since the late 1960s. The
unemploymentrate shows the most dramaticimprovementof the three
measuresin the 1990s, nearlyreturningto 1960s levels. By this common
21. AutorandDuggan(forthcoming).
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
99
Table 4. Changes in Nonemployment,by Reported Reason for Nonemployment,
1967-2000a
Percentagepoints
Reasonfor nonemployment
No work
available
Illness or
disability
Other
Total
Change1967-69 to 1999-2000
Nonemployment
Outof laborforce
Unemployment
1.2
0.6
0.6
1.7
1.7
0.0
1.8
1.7
0.0
4.7
4.0
0.6
Change1988-89 to 1999-2000
Nonemployment
Outof laborforce
Unemployment
-1.9
-0.4
-1.5
0.8
0.8
0.0
1.1
1.0
0.0
0.0
1.5
-1.5
Measure
Source: Authors' calculations based on March CPS data.
a. Details may not sum to totals because of rounding.
Figure 7. Changesin Unemployed,Discouraged,and NonemployedWorkers Since
1967, 1967-2000
Percent
10
Nonemployed
Other
nonparticipating
8
6
4
2
1970
1975
1980
1985
1990
1995
Source: Authors'calculationsbased on MarchCPS data.
a. Discouragedworkersare those who reportnot being employed because they were unable to find a job; they are not included
in the labor force.
100
BrookingsPapers on EconomicActivity,1:2002
measureof labormarketperformance,events have come full circle, and
one mightarguethatthe naturalrateof joblessnesshas returnedto previous low levels. Adding nonparticipantswho are discouragedworkers
changes the conclusion slightly, althoughthe figure also demonstrates
thattherehas been no reductionin discouragedworkerssince the 1980s.
to give totalnonemploymentchangesthe
Addingin othernonparticipants
interpretation
substantially.By this measuretherewas no improvementin
overalljoblessnessfromthe late 1980s to 2000, despitefallingunemploymentrates.In this sense, changesin unemploymentprovidea misleading
picture of changes in employmentopportunitiesand the likelihood of
findingwork.
The Incidence and Duration of Jobless Spells
The dataon full-yearnonemploymentsuggestthatthe concentrationof
unemploymentand nonemploymentincreased dramaticallyover the
periodcoveredby our data.Table 5 providesfurtherevidence, showing,
for variousperiods,the distributionsof spells of joblessnessduringa calendaryear. The trendtowardlong-termjoblessness is unmistakable.For
rate was 6.3 percent,
example, in the 1960s, when the nonparticipation
men who were jobless for the entireyear accountedfor 28.8 percentof
nonemployment.But by the end of the 1990s-when nonemployment
reached11 percent-full-year nonemploymentaccountedfor over half of
all joblessness. A similar patternholds for unemployment(table 6).
Althoughunemploymentratesin 1999-2000 were roughlycomparableto
those in the 1960s, the shareof unemploymentdue to shortspells (one to
thirteenweeks) fell by one-third,from30 percentto 20 percent.Individuals with more than six monthsof unemploymentaccountedfor about a
quarterof all unemploymentin the 1960s,but46 percentby the end of the
1990s. These shifts toward long-termjoblessness mean that particular
have much differentmeanrates of unemploymentand nonparticipation
ings todaythanin pastdecades.
To examinethe increasedimportanceof long spells more closely, we
use informationin the CPSto estimateboththe incidenceandthe duration
of jobless spells. Focusingfirston unemployment,we note thatthe rateof
unemploymentcan be decomposedinto the productof two components:
the probabilityof an individualenteringunemployment(the entryrate),
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
101
Table 5. Distributionof Nonemployment,1967-2000a
Percentof nonemployed
Numberof weeksa year
Period
1967-69
1971-72
1972-73
1975-76
1978-79
1982-83
1988-89
1991-92
1999-2000
2 orfewer
3 to 12
13 to 25
26 to 38
39 to 51
52
2.4
1.5
1.6
1.1
1.5
0.8
1.0
0.8
0.7
17.2
12.9
12.4
9.9
13.1
8.1
10.3
8.8
8.0
18.9
18.8
17.2
16.6
17.7
13.9
13.2
12.7
9.7
18.0
21.2
21.7
22.5
20.2
20.6
19.0
19.1
13.9
14.8
15.9
15.3
17.2
14.4
18.3
15.2
14.9
12.2
28.8
29.7
31.9
32.7
33.2
38.3
41.3
43.8
55.5
Source: Authors' calculations based on MarchCPS data.
a. Details may not sum to 100 because of rounding.
and the averagedurationof an unemploymentspell. Denote the instantaneous transitionratesfromemployment(e) andout of the laborforce (o)
to unemployment(u) at date t by Xeu(t) and XOu(t),respectively,and the
correspondingrates at which individualsleave unemploymentby Xue(t)
andXkj(t).Thenthe rateof changein the unemploymentrateis
(1)
du(t) =
dt
e(t)Xeu(t)
+ o(t)0ou(t)
- U(t)[Rue(t)
+
X0o(t)].
The steady-statefractionof weeks spentunemployed,[du(t)ldt= 0], correspondingto the entryandexit ratesat any given pointin time satisfies
Table 6. Distributionof Unemployment,1967-20O0
Percentof unemployed
Numberof weeksa year
Period
1967-69
1971-72
1972-73
1975-76
1978-79
1982-83
1988-89
1991-92
1999-2000
13 orfewer
14 to 26
27 to 39
40 to 49
50 to 52
30.3
20.4
20.6
17.8
27.0
13.7
22.6
16.9
20.4
42.3
40.6
39.9
31.2
34.2
28.9
33.5
31.7
33.4
14.9
21.5
21.0
22.8
19.0
21.7
18.5
20.6
18.2
8.9
10.9
10.4
14.5
11.8
17.9
14.8
16.4
14.6
3.6
6.6
8.2
13.7
8.0
17.7
10.5
14.3
13.4
Source: Authors' calculations based on MarchCPS data.
a. Details may not sum to 100 because of rounding.
102
BrookingsPapers on EconomicActivity,1:2002
()
u(t) = [1-uWWI
(2)
)
where
e (t)
X-(t) =
'u(t) =
Xue
X-eu(t)
+
o*
(t) X,u(t), and
(t) + X-u(t).
Here Xkis the rate at which individualsenter unemployment,being a
share-weightedaverageof entryratesfor personswho are employedand
those who are out of the laborforce. Similarly,X'(t) is the rateat which
individualsleave unemploymentby becomingemployedor by leavingthe
labor force. Since 1/X'(t) is the average durationcorrespondingto the
contemporaneousrateof exit from unemployment,and [1 - u*(t)]X.(t) is
the expectednumberof spells of unemploymentper year at the current
entry rate, equation2 has a naturalinterpretationin terms of entry and
duration.Growth in the steady-statefraction of the year spent unemployed can be decomposedinto growthin the probabilityof becoming
unemployed(entry)andthe averagedurationof unemploymentspells.
To implementthis frameworkempirically,we use two identitiesthat
correspondto equation 1 integratedover the year. The change in the
unemploymentratefromthe beginningto the end of yeart is
(4)
U,(t) - UO(t)
=
[1- U (t)]u () - U (0)u ('0,
where U1(r)is the unemploymentrate(measuredas a fractionof the population)at the end of yearX, UjQr)is the correspondingrateat the startof
the year, and U(t) is the averageunemploymentrateover the year.With
these definitions,Xu(t) andX' (t) are weightedaveragesof the instantaThe expected
neous transitionprobabilitiesto andfromunemployment.22
numberof spells of unemploymentover the yearis then
(5)
S(t) = Uo(t) + [1 - U
()r),
since spells are generatedeitherby startingthe year unemployed,UO(t),
or by becomingunemployedduringthe year, [1-U(r)]Xu(t). To estimate
the entryandexit parameters,we use the datafromthe CPS togetherwith
22. The weights in these weighted averages are (1 - u)(0) and u(O), respectively, where
e indexes weeks over the year.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
103
monthly data on aggregaterates to interpolatethe startingand ending
numbersfor each year. Solving equations4 and 5 gives our estimating
equationsfor unemploymenttransitionsas
(6)
()
S(t)-U0(t)
1-U(t)
and
(7)
(u
S(t)- U,()
The resultingestimatesareshownin the top panelof figure8 andin the
first two data columns of table 7. For unemployment,the key findingis
thatan increasein durationsaccountsfor the entiregrowthin unemployment over the 1967-2000 period.The entryrateinto unemploymentwas
actuallylower in 1999-2000 (0.7 percenta month)thanit was in 1967-69
(1.1 percenta month),whereasdurationsof unemploymentspells doubled
from2.1 to 4.2 months.Notice fromequation3 thata decliningincidence
of unemploymentspells can be caused eitherby a decline in the rate at
which individualslose their jobs, Xeu(t), or by a decline in the rate at
which nonparticipantsstartto look for work, XOu(t).
These contributions
arenot separatelyidentified,althoughit is likely thatthe o -* u transition
has declined substantiallyas nonparticipationhas become a permanent
laborforce state for largernumbersof men. In any case, we cannotconclude fromtable6 andfigure8 thatthe e -* u transitionhas declined,that
is, thatjobs have becomemore stable.
Accordingto figure8, untilthe recessionof 1991-92, cyclical fluctuations in unemploymentwere drivenby changesin boththe incidenceand
the durationof spells, with roughly equal weights on each component.
But rising incidence played a minor role in the recession of 1991-92,
while durationssoared. Indeed, unemploymentdurationsin 1993 were
virtuallythe sameas in the recessionyear 1983, whichwerethe highestin
all the yearsof our data,while the entryratewas about25 percentlower.
The ensuingdeclinein unemploymentover the remainderof the decadeis
drivenalmostentirelyby reducedprobabilitiesof becomingunemployed;
durationsof unemploymentremainedhigh. From these data it appears
that the main characteristicof the 1990s is that the historic correspondence between the incidence and the durationof unemploymentspells
BrookingsPapers on EconomicActivity,1:2002
104
Figure 8. EstimatedEntry Rates and Durationsfor Unemploymentand
Nonemployment,1967-2000
Unemploymenta
Monthsa year
Percenta month
2.2 -
Duration
(rightscale)
2.0 -
A"
5.0
/
1.8-
,*/13
{/
\~~~~~~~~~~~
04
1
45
1.61.4 --3.5
- 3.0
0.8
2.5
Entry rate
,'
(leftscale)
_ 2.0
0.6 _ _
l
l
l
l
I
l
Nonemploymentb
Duration
(rightscale)
2.2 -
4
,
2.0 -
1
1.8 -
,1
1.6 11.4 -(left
10
scale)
1.2
8
6
0.8
0.6 _
l
l
l
l
l
I
1970
1975
1980
1985
1990
1995
Source: Authors'calculationsbased on MarchCPS data.
a. Estimatedfrom the numberof spells using equations5, 6, and 7.
b. Based on incidence of nonemployment;see text for details.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
105
Table 7. Estimated Entry Rates and Durations of Unemploymentand
Nonemployment,1967-2000
Units as indicated
Unemploymenta
Nonemploymentb
Phase of
businesscycle
Entryrate
(percenta
month)
Duration
(months)
Entryrate
(percenta
month)
Duration
(months)
1967-69
1971-72
1972-73
Peak
Trough
Peak
1.1
1.5
1.4
2.1
3.2
2.9
1.0
1.2
1.1
6.7
8.6
8.5
1975-76
Trough
1.8
4.1
1.5
9.6
1978-79
1982-83
1988-89
1991-92
1999-2000
Peak
Trough
Peak
Trough
Peak
1.5
1.9
1.3
1.5
0.7
3.1
5.1
3.5
4.5
4.2
1.4
1.5
1.1
1.3
0.8
8.4
12.0
11.0
12.4
15.1
Period
Source: Authors' calculations based on March CPS data.
a. Estimatedfrom the numberof spells using equations 5, 6, and 7.
b. Based on incidence of nonemployment;see text for details.
came to an abruptend. Withfewerbutlongerspells, the populationdistributionof unemploymentis muchmoreconcentratedthanin earlieryears.
The last two columnsof table 7 andthe bottompanelof figure8 show
correspondingcalculationsfor the incidenceanddurationof nonemployment spells. In the case of nonemployment,the CPS does not provide
informationon the numberof spells in a calendaryear-separate spells of
arenot recorded-and so we use dataon the incidenceof
nonparticipation
nonemploymentover the year (thatis, the fractionof men with positive
weeks of nonemployment)to infer the entry rate.23These calculations
show thatthe contrastbetweenentryand durationis even more extreme
than in the case of unemployment.As with unemployment,the rate of
entryto nonemploymentis actuallylowerin 1999-2000 thanit was in the
1960s, but durationsshow a steady upwardtrend over the thirty-four
23. The fractionof individualswho experiencezero nonemployment(thatis, who are
employedfor the full year)is given by F(r) = Eo(r)exp[-12X*(r)],whereX*(t)is the average monthlynonemploymenthazardover the yearfor individualswho havenot yet entered
nonemployment,andEj(t) is the employmentrateat the startof the year.In general,X*(1t)
for the populationof
<X (t), whereX (t) is the averagerateof transitionto nonemployment
employedpeople.This will cause our estimatesof entryandexit ratesto be biaseddownward.We attemptedto assess the magnitudeandvariabilityin this bias with similarcalculations for unemployment,where the numberof spells is recorded.In that case the bias
variedlittle over time, lendingsome confidencethatthis methodshouldnot be too faroff.
106
BrookingsPapers on EconomicActivity,1:2002
yearscoveredby ourdata,withno sign of slowergrowthin the 1990s. By
the end of the decadethe averagedurationof nonemploymentspells was
over fifteenmonths,which is morethandoublethe lengthof spells in the
late 1960s. The averagedurationof spells rose by over fourmonthsfrom
the late 1980s to 1999-2000, reflectingthe increasingproportionof men
who have simplyquitthe laborforce.
Table 7 and figure8 painta clear picture.Althoughthe ratesof entry
into unemploymentand nonemploymenthave returnedto or even fallen
below levels experiencedduringthe late 1960s, the durationsof jobless
spells aremorethantwice as long at the end of the period.Indeed,jobless
spells were longer in 1999-2000 than at any previous cyclical low of
unemployment,andthey exceededthe averagedurationof spells over the
whole period of the data. It should be clear from these data that the
employmentpatternsof the late 1990s resemble other periods of low
unemployment-the late 1960s in particular-only in termsof the overall
rate of unemploymentand the rates at which individualsenterjoblessness. The durationsof spells arevery differentandvery muchlonger.For
the typicalworker,the occurrenceof a jobless spell is a fardifferentevent
thanit was in the past.
Unemployment,Nonemployment,and Wages
Our previous analysis of wage and employmentdata through 1989
foundthatthe patternsof changein unemploymentand nonemployment
variedsignificantlyacrossskill groups,as definedby percentileintervals
of the overall wage distribution.Figure 9 and table 8 summarizeour
results based on wage percentilegroupingsfor the period 1967-2000.
Table 8 recordschanges in unemployment,nonparticipation,and total
joblessness between 1967-69 and 1988-89 (the end of the data in our
1991 paper),between 1988-89 and 1999-2000, and over the full period
of ourdata.
All componentsof nonemploymentincreasedthe most for low-wage
workers,especiallybefore 1989. Overthe 1990s, nonparticipation
continued to rise while unemploymentratesdeclinedsharply.Reversingprevious trends,in the 1990s bothunemploymentandoverallnonemployment
fell the most for workersin the bottom 10 percentof the wage distribution. Otherlow-wage groupsalso experiencedlower unemploymentover
Figure 9. Unemployment,Nonparticipation,and Nonemployment,by Wage
PercentileGroup, 1967-2000
Unemployment
Percentof calendaryear
16
I tol /s0
-- l to 20
21 to 40 1_
-------41to 60 ''\
\
12
8-
Nonparticipation
20 15 10
S __-'
5,,
.
.
.~~~~~~~~~~~~~~~~~~~~~~~~
.....
...,,,,,,.,,
.....------.----
Nonemployment
35
3025 20
1970
1975
1980
Source: Authors'calculationsbased on MarchCPS data.
1985
1990
1995
108
BrookingsPapers on EconomicActivity,1:2002
Table 8. Changesin Unemployment,Nonparticipation,and Nonemployment,by
Wage Percentile Group, 1967-2000
Percentagepoints
Wagepercentilegroup
Period
I to 10
11 to 20
21 to 40
41 to 60
61 to 100
Unemployment
1967-69 to 1988-89
1988-89 to 1999-2000
1967-69 to 1999-2000
6.4
-4.6
1.8
4.9
-3.1
1.8
2.6
-1.6
1.1
1.5
-1.0
0.5
0.3
-0.5
-0.1
Nonparticipation
1967-69 to 1988-89
1988-89 to 1999-2000
1967-69to 1999-2000
10.9
0.8
11.7
7.2
2.4
9.5
2.9
2.6
5.5
1.3
1.3
2.6
0.0
0.9
0.9
Nonemployment
1967-69 to 1988-89
1988-89 to 1999-2000
1967-69 to 1999-2000
17.3
-3.8
13.5
12.1
-0.8
11.3
5.6
1.0
6.6
2.8
0.3
3.0
0.3
0.4
0.8
Full-yearnonemployment
1967-69 to 1988-89
1988-89 to 1999-2000
1967-69 to 1999-2000
10.2
1.7
12.0
6.3
2.7
8.9
2.9
2.4
5.4
1.6
1.3
2.9
0.5
0.9
1.4
Source: Authors' calculationsbased on MarchCPS data.
the decade,althoughnonemploymentwas largelyunchangedfor workers
above the bottom 10 percent,reflectingrising rates of laborforce withdrawal. Even with the sharp decline in unemploymentfor low-wage
workersin recentyears,however,for the full periodbothunemployment
andnonemploymentincreasedmost amongthe least skilled.Nonemployment rose by roughly 12 percentagepoints for the two lowest wage
groups,but by less than 1 percentagepoint for men above the 60th percentile of the wage distribution.For workersnearthe median(percentiles
41-60), unemploymentwas essentiallyunchangedover the period as a
whole, yet nonemploymentincreasedby 3 percentagepoints.
Cyclicalincreasesin joblessnessareknownto fall most heavilyon the
least skilled. Figure 10 comparesthe skill distributionsof cyclical and
secularchangesin nonemployment.Foreach of the wage intervalsshown
in table 8, the figure shows the averagecyclical change in jobless time
going into and out of four recessions(1970-71, 1975-76, 1982-83, and
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
109
Figure 10. Cyclical and Secular Changesin Nonemployment,by Wage Percentile
Group, 1967-2000
Percentage pointsa
EJAverage over recessionsb
1967-69 to 1988-89
* 1988-89 to 1999-2000
*
15 -
10
-
5
0
I
to 10
II to 20
21 to 40
41 to 60
61 to 100
Source: Authors' calculationsbased on MarchCPS data.
a. Change in the percentof weeks a year nonemployed.
b. Average change entering(troughminus peak) and exiting (peak minus trough)four recessions (1970-71, 1975-76, 1982-83,
and 1991-92).
1991-92),24the secular change between the cyclical peaks of 1967-69
and 1988-89 (the period covered in our 1991 paper), and the shorter
secular change over the more recent 1988-89 to 1999-2000 period.
Comparedwith business-cycleincreasesin nonemployment,the secular
changein nonemploymentfromthe late 1960s to the late 1980s was much
more skewed towardlow-skilled men, with virtuallyno impact on persons above the medianof the wage distribution.The secularmovement
over the recentperiodis morenearlyskill neutral,with the exceptionthat
nonemploymentfell significantlyfor men in the first decile of the wage
distribution.In this sense the 1990srepresenta smallreversalof declining
24. We measurethis by the averageof the changegoing into andout of each recessionary period. The periods are defined using the same year groupingsused in the tables:
1967-69, 1970-71, 1972-73, 1975-76, 1978-79, 1982-83, 1988-89, 1991-92, and
1999-2000.
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BrookingsPapers on EconomicActivity,1:2002
employment opportunitiesamong the least skilled. We relate these
changes to concomitantchanges in the distributionof wages below, but
firstwe takea briefdetourto explorean alternativeexplanationfor changingjobless ratesamongprime-agedmen, namely,the labormarketopportunitiesof theirwives.
Did Working Wives Shift Male Labor Supply?
It remainsour view that long-termchanges in male joblessness were
drivenby changesin labordemandthatdisproportionately
affectedlessskilled workers. These adverse demandconditions continued into the
1990s, althoughsomewhatmitigated,so thatnonemploymentcontinued
to rise even as measuredunemploymentwas falling. An alternative
explanation-with far differentwelfare implications-is that increased
wages andgreaterlaborforce participationof womenhave shiftedmen's
laborsupply:as the labormarketopportunitiesof wives improved,husbands chose to work less and household utility rose. On this view, at
least partof the long-termincreasein nonemploymentamongmen represents a welfare-improvingchangein householdlaborsupplydecisions.
Table9 andfigure11 explorethis issue. Table9 recordsmale earnings,
the percentageof householdswith a workingwife, averageearningsof
wives, and averagehouseholdincome for householdsin differentpercentilesof the malewage distribution.Amongless-skilledmen,wherethe
largest increasesin nonemploymentoccurred,the percentageof households with a workingwife actuallyfell over time. For these men, declining marriageratesoffset increasedlaborforce participationof women, so
that fewer low-wage men today reside with a working wife. For lessskilled men, averagehouseholdincome (which includes earningsof all
household members)increasedonly slightly from 1972-73 (when the
trendtowardrising inequalitybegan) to 1999-2000: averagehousehold
income rose by 11 percentin each of the percentileintervals1-10 and
21-40; in these groups, increases in the average earningsof working
wives by 40 percenthelpedto offset decliningmale earnings.In contrast,
the presenceof a workingwife is both higher and rising in households
above the 60th percentileof the male wage distribution,where men's
wages were rising and employmentrateswere stable.The shareof these
householdsin which a workingwife was presentincreasedby 9 percentage pointsafter1972-73 andby 12 percentagepointsfromthe 1960s, and
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
111
Table 9. HouseholdEarnings Characteristics,by Wage PercentileGroup, 1967-2000a
1996 dollarsa year,exceptwherenotedotherwise
Item
1st to I Othpercentile
Maleearnings
Percentof households
with workingwife
Wife's earningsb
Householdincome
21st to 40thpercentile
Maleearnings
Percentof households
with workingwife
Wife's earnings
Householdincome
61st to I 00thpercentile
Male earnings
Percentof households
with workingwife
Wife's earnings
Householdincome
All households
Maleearnings
Percentof households
with workingwife
Wife's earnings
Householdincome
1999-2000
1967-69
1972-73
1978-79
1988-89
11,134
11,354
11,479
8,625
9,584
38
3,050
22,965
36
3,392
26,366
36
3,766
28,140
31
3,975
25,952
29
4,770
29,194
23,746
25,213
23,579
21,522
21,507
49
5,132
35,394
48
5,773
39,137
48
6,133
40,770
46
7,410
40,773
40
8,091
43,431
44,213
49,375
47,147
51,449
59,843
41
4,786
54,456
45
5,918
61,722
50
7,219
61,861
55
11,537
71,937
54
14,993
88,507
31,230
34,121
32,686
33,193
36,789
44
4,713
42,263
45
5,548
47,538
48
6,388
48,560
49
8,895
52,821
45
10,854
61,541
Sources: Authors' calculationsbased on MarchCPS data and Economic Reportof the President, 2001.
a. Earningsand household income are deflated by the price index for personal consumptionexpenditure.Male's and wife's
earnings include income from wages, salary, and self-employment; household income includes all sources of earned and nonearned income of all household members.
b. Average of wife's earningsfor all males in the sample; the observationis zero when there is no working wife.
the averageearningsof these wives increasedby 153 percentbetween
1972-73 andthe end of the century.
It is difficult to square these facts with the view that long-term
increases in labor force withdrawalamong men have been driven by
improvedlabormarketopportunitiesfor theirwives. To settle the issue,
figure 11 shows changes in unemployment,nonparticipation,and fullyearnonemploymentfor men with andwithoutworkingwives. The clear
patternis thatrisingunemploymentandlaborforce withdrawalhave been
concentratedamongmen who do not have a workingwife. The contrastin
trendsis particularlystrikingfor nonparticipationand full-yearnonemployment,where men without a workingwife have steadily withdrawn
Figure 11. Male Nonemploymentand Nonparticipation,with and withouta Working
Wife, 1967-2000a
Nonemployment
Percentof calendaryear
20
Withoutworkingwife
15
Withworkingwife
Nonparticipation
12
F
10
86
42 -
Full-yearnonemployment
10
86 4 2
-
1970
1975
1980
1985
1990
1995
Source: Authors'calculationsbased on MarchCPS data.
a. To qualify as a workingwife, the wife must both live with her husbandand have workedat least one week duringthe previous year.
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
113
fromregularemployment.Fromthese datawe concludethata theorybuilt
on shifts in householdlaborsupplywill not go far in explainingchanges
in malejoblessness.
Wage Changes, Labor Supply, and Nonemployment
So far our discussionhas focused on changes in unemploymentand
nonemploymentover time. Figure 1 andtable 1 showedthatmanyof the
same patternsobservedwith regardto employmentand unemployment
hold for realwages. Inequalityin realwages grew significantlyfrom 1970
to 1990 across the full range of the wage distribution.Since 1990,
inequalityhas continuedto increaseat the top of the wage distribution,
but inequalityhas held steady or even narrowedslightly at the bottom:
both low-wage and middle-wage workers experienced real wage
increasesstartingaround1995. These increasesin real hourlywages representthe first significantgrowthin real wages for low- to middle-wage
males since the early 1970s. According to our earlier analysis, rising
wages for these groupsshouldlead to increasedemploymentrates,especially amongthe least skilled, for whom we concludedthatlaborsupply
elasticitieswere largest.
At a generallevel, trendsin nonemploymentby wage percentilegroup
(bottompanelof figure9) andtrendsin real wages for these same groups
(figure 1) reveal a similarpattern.In both cases low-wage workersfared
far worse thantheirmiddle-andhigh-wagecounterpartsfor muchof the
sampleperiod,and in both cases the divergencestops in the 1980s (after
roughly 1983 in the case of employment,and afterroughly 1989 in the
case of wages). Ourearlierpaperformalizedthis connection,arguingthat
declining rewardsto work provokedlabor supply responsesfrom lessskilled workers, who chose to work less. To what extent does the
demand-drivenexplanationwe stressedin our earlierpaper-that individuals respondto changingreal wage opportunities-help us to understand the changes since 1989 in employmentfor men in differentskill
categories?
Table 10 presentsestimatedpartiallabor supply elasticities obtained
fromcross-sectionaldatafor the years 1972-73 and 1988-89.25Ourestimates correspondclosely to those reportedin our earlier paper. They
25. To obtain these elasticities,we fit a quadraticfunctionusing averagewage and
employmentdataby percentilecategory.We reportthe slope at each percentile.
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BrookingsPapers on EconomicActivity,1:2002
Table 10. EstimatedLabor Supply Elasticitiesand Changesin Nonemploymentand
Wages, 1972-2000a
Unitsas indicated
Wagepercentilegroup
Partial
labor
labor
supply
elasticityb
Change inn
nonem
Changein
real lognoepymt
log
real
~~(percentagepoints)
hourlywage
(percent)
Predicted
Actual
1972-73 to 1988-89
1 to 10
11 to 20
21 to40
41to60
61 to lOO
Entiresample
0.287
0.217
0.170
0.126
0.048
n.a.
-24.8
-22.3
-16.5
-9.2
-0.2
-9.9
7.0
4.7
2.7
1.1
0.1
2.0
10.2
5.7
2.6
0.8
-0.1
2.2
1988-89 to 1999-2000
1 to 10
llto20
21 to40
41 to 60
61 to 100
Entiresample
0.287
0.217
0.170
0.126
0.048
n.a.
2.3
-0.4
-1.5
-1.2
7.0
2.4
-0.7
0.1
0.3
0.2
0.1
0.1
-3.8
-0.8
1.0
0.3
0.4
0.0
Source: Authors' calculationsbased on MarchCPS data.
a. Laborsupply elasticities and observed changes in real wages are used to predictthe change in nonemployment.
b. Elasticities in both panels are estimated using cross-sectional data on average wage and employment, by percentile, for
1972-73 and 1988-89.
show substantiallyhigherelasticitiesat low wages:the employmentrates
of less-skilled workersare more responsiveto wage changes. The top
panel of the table illustratesthe fact thatthe large wage declines (shown
in the second data column),togetherwith the estimatedelasticities,can
accountfor mostof the rise in nonemploymentfrom 1972-73 to 1988-89.
The bottompanelof the tableuses the same laborsupplyelasticitiesestimatedfrompre-1990data,togetherwith post-1989wage changes,to predict changesin nonemploymentduringthe 1990s. Thereis a reasonable
correspondencebetweenthe predictedandthe actualchanges:we predict
an improvementin employmentfor the lowest wage groupandsomewhat
worseningconditionsfor the othergroupsbelow the median.Yet we also
the improvementin employmentfor the least-skilledgroup.
underpredict
Althoughwages and employmentare obviously linked in the long run,
our interpretation
of the resultsis thatthe laborsupplymodel is less successful in predictingthe dynamicsof employmentandwage changesover
ChinhuiJuhn, KevinM. Murphy,and RobertH. Topel
115
a relativelyshortperiod.Notice also thatemploymentgains precededthe
recoveryin wages among the least skilled, which is inconsistentwith a
purelaborsupplyexplanationof changingemploymentrates.
Conclusion
We have examinedunemploymentandnonemploymentamongprimeaged males in the United States using thirty-fouryears of data on labor
marketoutcomes. Although recent unemploymentrates have fallen to
rates
levels reminiscentof the 1960s, we findthatrisingnonparticipation
have offset reductionsin unemployment,leaving nonemploymentrates
unchanged.The rise in nonparticipationappearsto be due to both an
expansionof the disabilitybenefits program-as previous researchhas
argued-and continuedlow levels of real wages of less-skilledmen during the 1990s.
in the
Comparedwith earlierdecades,the increasein nonparticipation
skill
groups.Employmentratesof
1990s is moreevenly distributedacross
the least skilled rose the most, even as their wages lagged behindother
groupsfor muchof the decade.This suggeststhatrisinginequality,which
characterizedlabor marketsin the 1970s and 1980s, may have run its
course.
Is the Americanlabormarkettoday fundamentallydifferentfromthat
of the 1960s?Despite the comparabilityof unemploymentratesbetween
the late 1960s and the late 1990s, the changingcompositionof nonemand from part-year
ployment-from unemploymentto nonparticipation,
to full-yearnonemployment-suggeststhatthe combinationof low wages
and the availabilityof nonworkalternativeshas madeout-of-workmales
increasinglyless likely to enter new jobs. From this perspective,our
assessmentof the labormarketfor less-skilledmen is rathergrim.
Ourearlierwork concludedthatthe naturalrate of unemployment,or
of nonemployment,is not a constanttowardwhich the economy gravitates over time. Rather,it varieswith labormarketconditionsin a manner
consistentwith the original formulationof EdmundPhelps.26Over the
long term, the naturalrate of nonemploymenthas increased because
26. Phelps(1974).
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BrookingsPapers on EconomicActivity,1:2002
changingpatternsof labordemandhave reducedthe returnsto work for
less-skilledmen. In this setting,the low unemploymentratesof the latter
half of the 1990s have a far differentinterpretation
thancomparablerates
of the past. By the end of the 1990s, an importantproportionof lessskilled men had withdrawnfrom the laborforce for demand-related
reasons. Thatthey arenot countedamongthose seekingworkis not a sign of
strengthin currentlabormarketconditions.