Crowding-Out Effect of Publicly Provided Childcare:
Why Maternal Employment Did Not Increase

ISS Discussion Paper Series F-177
University of Tokyo
Yukiko Asai, Ryo Kambayashi and Shintaro Yamaguchi

Crowding-Out Effect of Publicly Provided Childcare:
Why Maternal Employment Did Not Increase∗
Yukiko Asai†

Ryo Kambayashi‡

Shintaro Yamaguchi§

July 6, 2015

Abstract
We estimate the causal effects of childcare availability on the maternal employment rateusing prefecture panel data constructed from the Japanese quinquennial census 1990-2010. We
find that childcare availability did not increase maternal employment due to the crowding-out
effects. Namely, families substituted accredited childcare for informal care by grandparents.
We also find evidence that more and more families do not live with grandparents who used totake care of grandchildren, as the availability of accredited childcare increases.

Keywords: childcare; female labor supply; maternal employment; nuclear family;
three-generation family
JEL Codes: J13, J21, J22

1

Introduction

Maternal employment hinges on quality, affordability, and availability of childcare. In countries
where the childcare market is regulated and heavily subsidized, high quality and affordable childcare service is provided, but the supply often falls short of the demand. Expecting that supporting∗ This is an updated version of the paper entitled as “Childcare Availability, Household Structure, and Maternal
Employment” (forthcoming in the Journal of the Japanese and International Economies.) The authors thank the support
from Joint Usage and Research Center, Institute of Economic Research, Hitotsubashi University and are permitted touse Employment Structure Survey through KAKENHI (J130100386). Yamaguchi gratefully acknowledge the financial
supports from Zengin Foundation for Studies on Economics and Finance.
† JSPS Postdoctoral Fellow at Institute of Social Science, the University of Tokyo‡ Institute of Economic Research, Hitotsubashi University§ Corresponding author. Department of Economics, McMaster University, 1280 Main St. W., Hamilton, ON.
Canada L8S 4M4. Email: yamtaro@mcmaster.ca.

1

working mothers will raise the low fertility rate, the Japanese government has implemented a series of pro-family policies, including an expansion of capacity of accredited childcare centers since
early 1990s. In this paper, we estimate the causal effects of childcare availability on the maternal
employment using the prefecture panel data from the Japanese quinquennial census 1990-2010.
Even though mother’s employment is a prerequisite for enrollment in an accredited childcare
center in Japan, whether childcare availability significantly increases maternal employment is not
immediately obvious. If working mothers substitute the accredited childcare service for an informal childcare arrangement such as the one provided by grandparents, the maternal employment
rate will remain unchanged. Indeed, Fitzpatrick (2010) and Havnes and Mogstad (2011) find thatan expansion of childcare services did not raise the maternal employment rate in the U.S. and Norway, respectively, because the new childcare mostly crowded out informal childcare arrangements.
Most previous studies on Japan that estimate the effect of childcare availability on female labormarket outcomes rely on the cross-sectional variation between prefectures.12 Examples include,
but are not limited to, Shigeno and Ookusa (1999), Higuchi, Matsuura, and Sato (2007), Unayama(2011), and Abe (2013). While the reported positive correlation is suggestive, it may not necessarily imply the causal effect of childcare availability on the female labor market outcomes. Maternal
labor supply is affected not only by childcare availability, but also by the local industry structure, economic conditions, commuting time, traditional family values, and preference for women
working. Because traditional family values and preference for women working are unobserved
and hard to control for, the observed positive correlation between childcare availability and maternal employment may be driven by the unobserved characteristics, rather than indicating a causal
relationship.
This paper departs from the previous studies on Japan’s childcare employment effects by controlling for unobserved prefecture fixed effects. Using the ratio of childcare capacity to populationof children under 6 as a proxy for childcare availability, we regress the maternal employment rateon the childcare availability index along with year and prefecture dummy variables as well as otherobserved household and prefecture characteristics. Because the prefecture fixed effects are controlled, the causal effect of childcare availability is identified by within-prefecture variation in thegrowth of childcare availability. This approach provides more credible evidence for the causal effects than the previous approach that relies on the cross-sectional variations alone. Our approach
1 A similar approach is also taken by Brilli, Del Boca, and Pronzato (2013) for the study on Italy.
2 An exception is Lee and Lee (2014). Using aggregate data at the country level on childcare availability, the female

labor force participation rate, and the fertility rate from 1971 to 2009, they try to establish Granger-causality using atime-series econometrics technique. Our identification strategy is very different from Lee and Lee (2014) in that weuse variations in the growth of childcare availability across prefectures. Moreover, Lee and Lee (2014) do not include
the household structure or the nuclear household share in their VAR model. Our analysis shows that the household
structure strongly affects the maternal employment rate and has changed over time significantly. Omitting this variablein the VAR model might affect their estimation results.

2

is essentially the difference-in-differences estimator, which is adopted by Berlinski and Galiani(2007), Baker, Gruber, and Milligan (2008), Lefebvre and Merrigan (2008), Havnes and Mogstad(2011), and Nollenberger and Rodriguez-Planas (2015).3
Our results indicated that, contrary to popular belief, childcare availability is uncorrelatedwith the maternal employment rate when prefecture fixed effects are controlled. Because they
are strongly and positively correlated without prefecture fixed effects, this result implies that theestimates identified by cross-sectional variations alone are upward-biased. A possible explanation
for the uncorrelatedness is substitution of the accredited childcare service for the informal childcare arrangements. Because the childcare provided by grandparents is the most common formof informal childcare arrangement in Japan, we analyze nuclear households and three-generation
households separately. This analysis enables us to shed light on the role of informal childcare
provided by the grandparents in determining the maternal employment rate. We find that, evenwhen prefecture fixed effects are controlled for, childcare availability is strongly and positively
correlated with the maternal employment rate for nuclear households. However, they are uncorrelated for three-generation households when prefecture fixed effects are controlled. The results
suggest that the household structure is the key to understanding the relationship between childcare
availability and maternal employment.
Although we do not observe the household’s choice of childcare mode in the census, households seem to have substituted the accredited childcare service for informal care by grandparents.
The key factor behind this substitution is a change in the structure of households. Namely, an
increasing number of Japan’s households do not include grandparents any more. Our results froma fixed-effect regression indicate that the rise of childcare availability induced the shift from threegeneration households to nuclear households for households that included children under 6. Interestingly, we did not find a significant effect on households in which the youngest child is aged
6-14, which suggests that our results for households with children under 6 is not a mere correlation,
but a causal relationship.
Fitzpatrick (2010) and Havnes and Mogstad (2011) find a null effect of childcare expansion
on maternal employment in the U.S. and Norway because it crowded out informal childcare arrangements. In contrast, ? find a positive effect of childcare expansion in Spain. They argue that
crowding out did not occur in Spain, because the female labor force participation rate was low, thetraditional family values were rooted, and the childcare supply was insufficient. The labor market
institutions and circumstances surrounding childcare in Japan are more similar to Spain, but our
results are more in line with those for the U.S. and Norway, which may seem counterintuitive. The
3 An alternative approach to avoid an endogeneity bias is based on the Regression Discontinuity Designs.

Gelbach (2002), Fitzpatrick (2010), and Goux and Maurin (2010) exploit an age-dependent eligibility rule for childcare
enrollment and identify the causal effect.

3

key to understanding the different results between Japan and Spain is the household structure. In
Spain, the share of three-generation households was much lower than that in Japan, and hence, theinformal childcare by grandparents was uncommon and not crowded out.
Our results for Japan suggest that neither the low female labor force participation rate nor the
traditional family value themselves are the determinants of a positive causal effect of childcare
availability on maternal employment. In Japan, the traditional family values require the eldest son
to live with his parents. Even though the traditional family values do not encourage mothers towork outside the home, this requirement eventually makes wives’ labor force participation rate
high due to the availability of childcare by grandparents. Then, as the traditional family values
became less influential, more and more households did not live with grandparents, which in turnlowered the maternal employment rate. This negative effect offset the effect of improved childcare
availability during the same period. Note that this offsetting effect was more strongly pronounced
in small prefectures where the traditional family value is more rooted. Our analysis indicatesthat the prevalence of informal childcare is the key determinant of whether maternal employment
increases with childcare availability or not.
We address two potential threats to the identification of the causal effects of childcare availability. The first threat comes from households moving between prefectures for better childcare
availability on maternal employment. Our evidence from Japanese Panel Survey of Consumers
indicates that moving for childcare is very rare: at most 1% of households with children under 6
moved from other prefecture for the reason of better childcare availability at the most recent move.
The second threat is prefecture-level shocks such as changes in local labor market conditions andpro-family policies that coincide with the changes of childcare availability. We address this issue
by examining the employment rate of mothers whose youngest child is between 6 and 14 because
they should not be directly affected by childcare expansion. The result indicates that their employment rate is uncorrelated with childcare availability. We conclude that these two potential threats
to identification are negligible.
We examine how effects of childcare availability changed from the 1990’s to the 2000’s andfind that the employment effects turned into positive in the 2000’s. The regression result indicates that the share of nuclear household still increases with childcare availability, but the effectbecame substantially weaker. This implies that the crowding-out effect of accredited childcare became weaker. Consequently, in the 2000’s, the maternal employment rate increased with childcare
availability.
The rest of the paper is structured as follows. In Section 2 we describe the institutional background and data. Section 3 lays out the econometric model. Section 4 presents our estimation
results for the effects on maternal employment. Section 5 shows the results for the effects on thehousehold structure. We discuss how the effects of childcare availability changed from the 1990’s
4

to the 2000’s in Section 6. We conclude in Section 7.

2

Background and Data

2.1

Childcare Policies in Japan

Childcare service is strictly regulated in Japan. For a childcare center to be accredited and subsidized by the national and municipal governments it must satisfy the criteria for capacity, area
and the number of teachers per pupil, etc. Because accredited childcare centers are subsidized, theaverage monthly fee for a child is low, at 25,556 JPY (about 250 USD),4 although different fees
are charged depending on regions and the household income. Non-accredited childcare centers satisfy lower quality standards than those for accredited childcare centers. While some municipality
governments subsidize non-accredited centers, the national government does not subsidize them,which makes their fee higher. Their average monthly fee for a child age 0 is 46,330 JPY (about
460 USD) and that for child age 5 is 34,161 JPY (about 340 USD).5
According to the Comprehensive Survey of Living Conditions, in the period from 1998 to2010 92-93% of childcare centers were accredited and the remaining 8% were non-accredited.6
Because non-accredited childcare centers are not subsidized and of lower quality, they are uncommon except for the large prefectures such as Tokyo and Kanagawa where the supply of accredited
childcare is far exceeded by the demand. For example, the share of non-accredited childcare centers is 10% in Tokyo and 16% in Kanagawa, while it is nearly zero in small prefectures including
Gunma, Toyama, Fukui, Yamanashi, Nagano, Wakayama, and Tottori.
Thanks to regulations, the quality is high and the price is affordable, but the accredited childcare
service is not available for all households in need. Insufficient supply of the accredited childcare
has been considered an obstacle for mothers’ entering the labor market and a cause of the low
fertility rate in Japan. The Japanese government implemented a series of policies for raising thefertility rate since the mid 90’s. The Angel Plan (1994-1998) and the New Angel Plan (19992003) are agreements between ministers and intended to support working mothers. They include
an expansion of childcare capacity, extension of childcare service hours including weekends andholidays, and subsidies to promote the take-up of parental leave and shorter working hours. However, these plans do not have legally binding power, and consequently, they failed to increase thecapacity of accredited childcare. Indeed, this capacity decreased by 10% from 1990 to 2000, eventhough the childcare capacity per child population slightly increased due to the low fertility rate.
4 See Table 7 on page 8 of Ministry of Health and Wealth (2009)
5 See Table 12 on page 14 of Ministry of Health and Wealth (2009)
6 See Table 40 in Volume 2 for 1998, Table 24 in Volume 3 for 2001, and Table 18 in Volume 3 for 2004, 2007, and

2010.

5

In 2003, the Japanese government enacted the Basic Act for Measures to Cope with Society with
Declining Birthrate, which is legally binding. Since then, the capacity of the accredited childcare
centers increased by a significant 12.2% from 2000 to 2010.

2.2

Measuring Childcare Availability

Our key explanatory variable is an index of childcare availability at the prefecture level, which
is defined as the ratio of the capacity of the accredited childcare to the population of children
under 6. As mentioned above, the share of accredited childcare centers has been at 92-93%, andhence, the proposed index roughly measures overall availability of center-based childcare. In thispaper, childcare capacity and enrollment statistics cover accredited childcare centers only. We draw
childcare capacity data from the annual Report on Social Welfare Administration and Services7 andchild population data from the Japanese quinquennial census 1990-2010. We emphasize that ourmeasure of childcare availability is based on capacity, not on enrollment. Because capacity is a
supply side factor and does not pick up households’ willingness to use, we interpret our index as ameasure of childcare supply relative to child population.
An alternative measure of the childcare availability is the number (or proportion) of children on
the waiting list for accredited childcare.8 However, there are two reasons against using the waiting
list as a good measure. First, in 2003, the government changed the definition of waitlisted children,which makes the statistics before and after 2003 not comparable. Second, although the size of theexcess demand can be useful for measuring availability, the waiting list does not measure the excess
demand. If parents expect a very low chance of getting accepted by an accredited childcare center,they are unlikely to submit an application to avoid a time cost of the application. For example, cityof Yokohama declared that no child was waitlisted in April 2013, but the application increased by
4,114 or 8% in the following year.9 This anecdote suggests that the actual excess demand for theaccredited childcare is greater than the number of waitlisted children.
As mentioned in the introduction, the causal effect of childcare availability on maternal employment is identified by the variation in the growth (instead of the level) of the childcare availability index across prefectures. It is worth describing how the variation was generated during thesample period. Since early 1990’s, a series of policies including The Angel Plan, the New Angel
7 Childcare capacity data are also available from Survey on Social Welfare, but data are not comparable before andafter 2007.
8 Yet another possible measure of childcare availability is proposed by Unayama (2011) who studies the relationshipbetween childcare availability and job continuity at marriage. His index is the ratio of childcare capacity to the numberof women age 20-44. His rationale is that the number of women age 20-44 measures the potential number of children.
Our index based on the actual number of children directly measures the current availability, which we think is more
relevant for decisions by mothers who currently have children under 6. Despite the differences in concept, the results
do not change regardless of whichever measure is used. See Tables for 7 and 8 for evidence.
9 See Yokohama (2014) for their press release.

6Plan, and the Basic Act for Measures to Cope with Society with Declining Birthrate promoted raising the childcare capacity across the country. In large prefectures such as Tokyo, the capacity wasincreased as intended by the policies, but in small prefectures it remained almost constant, because
the child population has been rapidly decreasing. Note that childcare availability improved even insmall prefectures, since the child population decreased, but the capacity remained unchanged. The
childcare capacity does not decrease proportionately to the child population for two institutional
reasons. First, there is a regulation that the minimum capacity of an accredited childcare be 60
children.10 This minimum capacity requirement directly prevents the operators of the childcare
centers from downsizing. Second, the Japanese dismissal law strongly protects workers, which
prevents childcare centers from firing them to reduce their capacity.11

2.3

Household and Prefecture Characteristics

We draw our data on household and prefecture characteristics from the Japanese quinquennial
census from 1990 to 2010.12 Our analysis focuses on households with two parents and children
under age 6. The share of two-parent households among all households with children under 6 is
as high as 97-98% from 1995 to 2010, according to the Census. These households are categorized
by the census into three groups. The first household type is a nuclear household in which only
two parents and children reside. The second household type is a three-generation household thatconsists of two parents, children, and grandparents. The third household type is other householdthat consists of two parents, children, and other adults such as relatives. This last type is only about
0.02% of all two-parent household with children under 6.
In the census, employment is identified by whether an individual performed paid work or notfrom September 24th to 30th of the year. If an individual did not work but was on leave, she/heis considered employed by the census. The fraction of employed individuals and the average age
are reported for the husband and wife at the prefecture level. As a measure of local economic
condition, we construct an unemployment rate for individuals age 15 and over at the prefecturelevel.

2.4

Summary Statistics

Table 1 reports the mean of each variable by year and household type, using the number of households as a weight. The unit of observation is prefecture and there are 47 of them. The population is
two-parent households with children under 6. Note that the means are different between the house10 Nihon-Jidou-Mondai-Chousakai (1978) documents how the minimum capacity is determined.
11 See p. 512 in Yamakawa (2007)
12 All of our data from the census 1990-2010 are publicly available and downloadable from www.e-stat.go.jp.

7hold types, because the numbers of nuclear households and three-generation households in a given
prefecture are different. Our childcare availability index, which is defined as the ratio of childcare
capacity to the population of children under 6, steadily increased over time from 0.248 in 1990 to
0.338 in 2010, on average over all types of households. The average of the capacity-population
ratio for nuclear households is lower than that for three-generation households, which reflects thefact that the share of nuclear households is higher in metropolitan areas where childcare services
are less available.
The employment rate of mothers with children under 6 rose from 34.1% in 1990 to 41.2% in
2010; a remarkable employment growth given the long-term recession. This increase is largely
driven by the growing employment rate of mothers in nuclear households: it increased from 27.4%
in 1990 to 39.6% in 2010. Although the employment rate of mothers in three-generation households is as high as 51.1% in 1990, it decreased to 47.8% in 1995 and gradually returned to the1990 level by 2010.
Husbands’ employment rate decreased from 98.8% in 1990 to 94.1% in 2010. In contrast to
women, we find little difference in male employment rate across household types.
Mothers’ average age increased from 31.580 in 1990 to 33.720 in 2010 while that of husbands
increased from 34.454 in 1990 to 35.632 in 2010. The rise in parents’ age reflects the fact that men
and women delayed childbirth in the past few decades. We find little difference in parents’ age
across household types.
The average local unemployment rate rose from 3% in 1990 to 6.4% in 2010.13 Despite nuclear
and three-generation households tending to reside in different regions, we find little difference inlocal unemployment rates faced by the nuclear and three-generation households.
The share of nuclear households dramatically increased from 71.2% in 1990 to 86.5% in 2010.
The number of two-parent households decreased from 5.5 million in 1990 to 4.5 million in 2010.
Because the share of two-parent households among all households with children under 6 is stable
at 97-98%, this trend reflects the declining fertility rate.

3

Econometric Model

We analyze the causal effects of childcare availability using a regression model. Consider thefollowing econometric model
Ypt = β0 + β1CCA pt + β2 X pt + θ p + ξt + ε pt ,(1)

13 Note that the highest national unemployment rate during this period is 5.5%. The average of local unemployment

rate in 2005 and 2010 exceed this number, because we weight by the number of household with children under 6.

8

Table 1: Characteristics of Prefectures and Two-Parent Households with Children Under 6

All HH
Cap./U6-Pop.
% Mothers EmployedAge
Age of Husband% Husband Empl.
Unemp. Rate
% Nuclear HH
Nuclear HH
Cap./U6-Pop.
% Mothers EmployedAge
Age of Husband% Husband Empl.
Unemp. Rate
3-Generation HH
Cap./U6-Pop.
% Mothers EmployedAge
Age of Husband% Husband Empl.
Unemp. Rate

1990

1995

2000

2005

2010

0.248
0.341
31.58034.454
0.988
0.030
0.712

0.265
0.329
31.75734.517
0.985
0.043
0.752

0.269
0.334
32.14234.564
0.976
0.048
0.804

0.302
0.372
32.88334.953
0.965
0.060
0.837

0.338
0.412
33.72035.632
0.941
0.064
0.865

0.241
0.274
31.56034.373
0.988
0.031

0.258
0.280
31.63234.313
0.984
0.044

0.263
0.300
32.04934.382
0.976
0.048

0.296
0.347
32.88634.879
0.965
0.060

0.332
0.396
33.76335.614
0.941
0.064

0.266
0.511
31.62834.652
0.990
0.028

0.285
0.478
32.14735.146
0.986
0.039

0.293
0.476
32.53735.327
0.975
0.044

0.331
0.499
32.88435.346
0.968
0.057

0.376
0.517
33.45835.7500.945
0.064

Source: Census 1990-2010.
Note: Unit of observation is the prefecture, and 47 prefectures exist. The population is two-parent
households with children under 6. The number of households is used as weight.

9where Ypt is the maternal employment rate in prefecture p in year t, CCA pt is our measure ofchildcare availability given by the ratio of capacity to child population, X pt is a vector of observed
characteristics of household and prefecture, θ p is prefecture fixed effects, ξt is year fixed effects,
and ε pt is an error term that is uncorrelated with other variables. The fixed effects θ p and ξt areallowed to be correlated with other variables including CCA pt . The prefecture fixed effects capture unobserved household and prefecture characteristics including the traditional family value
and preference for maternal employment. The year fixed effects capture factors that affect maternal employment equally across prefecture. For example, the effects of a nation-wide secular risein female wages and changes in parental leave and other labor legislation14 by the national government are included in the year fixed effects. In all our regression models we use the number ofhouseholds as weights and report standard errors clustered at the prefecture level.
Although we could measure childcare availability at the municipality level, we choose to measure it at the prefecture level for two reasons. First, the use of the prefecture level data avoids
a potential endogeneity bias due to inter-prefecture migration. Our identification assumption forEquation (1) is that the error term is uncorrelated with the growth of childcare availability CCA pt
at the prefecture level. A possible threat to this assumption is migration for the reason of betterchildcare availability. In Section 4.3.1, we show evidence that inter-prefectural migration for betterchildcare availability is extremely rare. This exogeneity assumption may be questionable if childcare availability is measured at the municipality level, because intra-prefecture migration is morecommon.
The second reason is that the use of the prefecture level data minimizes a potential attenuation bias due to the measurement error. This is because there are fewer observations in eachmunicipality-year cell than in each prefecture-year cell. Variables in Equation (1) should be population means, but in practice we have to use sample means as proxies. When the municipality level
data is used, the population mean may be imprecisely estimated due to the small sample size.
A potential downside of using observations at the prefecture level, rather than the municipality
or individual level, is a lack of variation in the explanatory variables. This lack in variation may
result in unstable estimates or large standard errors, but our standard errors are small as shown
below.
14 See Asai (2015) and Yamaguchi (2014) for the employment effects of parental leave legislation.

10

4Effects on Maternal Employment

4.1

All Household Types

4.1.1

Overall Effects

Table 2 reports the regression results. All standard errors are clustered at the prefecture level.
The population for columns (1) through (3) is mothers of children under 6 in all household types
including nuclear, three-generation, and other households.
Without any control, the coefficient for the childcare availability index is 0.686 (column 1).
We then include in the regression the average age of mothers, that of fathers, the employment rateof fathers, the local unemployment rate at the prefecture level, and year dummies. When thesecontrol variables are included, the coefficient becomes slightly smaller at 0.567, but still strongly
positive and significant (column 2). This is consistent with the findings of previous studies on
Japan mentioned above. Note that the positive correlation may not be interpreted as a causal
effect of childcare availability on the maternal employment rate, because it may be driven byunobserved prefecture characteristics such as the traditional family values, preference for maternal
work, industry structure, commuting time, etc.
We control for these prefecture fixed effects by adding prefecture dummies. The coefficientfor the childcare availability index in this specification is identified by the variations in the growth(instead of the level) of employment rates and childcare availability across prefecture. The result incolumn (3) is surprising: the estimated coefficient is -0.147, although it is statistically insignificant.
The comparison of columns (2) and (3) warns us that the omitted variable bias is severe enoughto change the sign of the estimates. This result indicates that the positive correlation between
childcare availability and maternal employment is not a causal relationship, but driven by theprefecture characteristics.
4.1.2

Uncovering The Mechanism

Why doesn’t the maternal employment rate respond to childcare availability? Our finding is similar to those of Fitzpatrick (2010) and Havnes and Mogstad (2011). They find that an expansionof childcare did not increase maternal employment in the U.S. and Norway, because it crowded
out informal childcare arrangements by unlicensed caregivers such as babysitters. In Japan, childcare by grandparents is the most prevalent form of informal childcare arrangement. Longitudinal
Survey of Newborns in the 21st Century asks who is the main caregiver for children under 3.15
According to the survey, among children who are looked after by someone else but their parents,
15 The survey collects the information on the main caregiver for children age 3 and above, but it is not comparable

to the information for children under 3.

11

73.3% of the children are enrolled in childcare centers, 25.9% of the children are looked afterby grandparents, and only 0.9% of the children are under other informal childcare arrangements
including babysitters and nannies.
Hence, we suspect that households substitute the accredited childcare for the care by grandparents. Unfortunately, we do not observe the households’ choice of childcare mode in the census, but
we can still shed light on the relationship among the accredited childcare, the care by grandparents,and maternal employment. Evidence from the Comprehensive Survey of Living Conditions 2010
indicates that the share of children who are looked after by grandparents16 tends to be higher in
prefectures where the share of three-generation households is high. The correlation coefficient for
the two is as high as 0.80, which suggests that children in three-generation households are likely
to be looked after by their grandparents.
Childcare availability may affect families’ decisions on co-residence with grandparents. Grandparents can domestically produce childcare service, which gives rise to the surplus from forming an
extended family household. However, the expansion of publicly-funded childcare service increases
its availability and lowers the effective price of childcare service in the market, which reduced thesurplus for an extended family. Because young couples value their independence and privacy, they
may choose not to live with grandparents when public childcare service is available. Hence, weview co-residence with grandparents is a mediator. Childcare availability affect maternal employment not only directly, but also indirectly through affecting co-residence with grandparents.
To evaluate the direct effect of childcare availability on maternal employment, we include theshare of nuclear households in the regression. This approach implies that the household structure is taken as exogenous, but women whose labor force attachment is strong might choose toform a three-generation household, rather than a nuclear household, because informal childcare
may be available in three-generation households. Sasaki (2002) addresses this issue and estimatesthe effect of co-residence with the parents or in-laws on the wife’s labor force participation. He
instruments co-residence by whether the husband is the eldest son or not, because the traditional
Japanese family value obligates the eldest son to reside with his parents. Sasaki (2002) finds that
his OLS estimate for the effect of co-residence is not significantly different from the IV estimate.
This implies that co-residence can be taken as exogenous to the wife’s labor force participation,
which justifies our approach.
The model (4) includes the share of nuclear households in addition to all the control variables ofmodel (3). The coefficient for the childcare availability index is 0.117 and statistically significant
at the 10% level. Remember that the childcare availability index grew by 0.090 from 1990 to2010. This implies that the maternal employment rate would have been increased by 1 percentage
point (= 0.117 * 0.090) if the share of nuclear households did not change. Given that the maternal
16 See Table 18 in Volume 3.

12

employment rate grew from 0.341 in 1990 to 0.412 in 2010, the growth of childcare availability
would account for about 15% of the growth of maternal employment rate during the two decade
when the nuclear household share was kept at the 1990 level.
Once the share of nuclear households is included in the regression, the sign of the coefficientchanges from negative (column 3) to positive (column 4), which implies that the changes of thechildcare availability index and the nuclear household share are positively correlated. If the rise in
the nuclear household share is a result of better childcare availability, further expansions of childcare will lead only to more nuclear households, without necessarily raising maternal employment.
If the increase in the share of nuclear households occurs independently from a change in childcare
availability, childcare expansion can raise maternal employment as long as the pace of expansion
is fast enough to exceed the negative effect from the increase of the nuclear household share. We
address this issue in Section 5.
We also include an interaction term of the childcare availability index and the nuclear household share (column 5) to examine how the treatment effects vary across prefectures with the nuclear
household share. The coefficients are insignificant, but the positive sign for the interaction termsuggests that the treatment effect is stronger for prefectures where the nuclear household share is
high.

4.2

Nuclear and Three-Generation Households

We report the regression results for mothers from nuclear households in columns 6-8 of Table 2.
The size of the coefficient decreases as we add more control variables, which suggests the estimates
are upward biased when control variables and prefecture fixed effects are omitted. Nevertheless,the coefficient remains positive and significant 0.322 (column 8) for nuclear households even when
the prefecture fixed effects are controlled.17 The estimate implies that the maternal employment
rate for nuclear families increased by 2.90 percentage points (= 0.322 * 0.090) from 1990 to 2010,
which accounts for about a quarter of the growth of the maternal employment rate for nuclear
families (0.122 = 0.396 - 0.274).
Columns 9-11 show the regression results for mothers from three-generation households thatconsist of grandparents, parents, and children.18 The estimated coefficients are positive when noor some observed control variables and year dummies are included (columns 9 and 10), but thecoefficient becomes negative once the prefecture fixed effects are added (column 11). The regression results suggest that the employment of mothers from three-generation households does not
respond to changes in childcare availability: the estimated treatment effect is -0.243 and statisti17 This result is robust to the inclusion of prefecture-specific trends in the maternal employment rate. See an earlier

version of our paper, Asai, Kambayashi, and Yamaguchi (2014).
18 Other individuals such as relatives may or may not live together.

13

cally insignificant (column 11). One possible explanation for this result is that three-generation
households do not enroll their children in the accredited childcare, because the informal care by
grandparents is already available. Another possible explanation is that three-generation households
substitute the accredited childcare for the grandparents’ care. Either way, childcare availability
does not affect the employment rate of mothers from three-generation households.

4.3

Robustness

Although the regression with prefecture dummies controls for prefecture fixed effects, there might
be a simultaneity bias. Namely, there may be a correlation between childcare availability andmothers’ unobserved willingness to work even when the time-invariant household and prefecture
characteristics are controlled.
One possible source of the simultaneity bias is households moving to a prefecture where childcare service is more available. Another possible source is prefecture-level changes in policies andlabor demand shocks that affect not only mothers of children under 6, but also similar demographic
groups. For example, pro-family policies unrelated to childcare at the prefecture level might increase the employment of mothers of older children. In the following, we show that these two
possibilities are negligible.
4.3.1

Inter-Prefecture Migration

Popular narrative says that obtaining a spot in an accredited childcare center is extremely hard inTokyo and that some people even move to other municipality for childcare. For example, in thearticle from The New York Times, Tabuchi (2013) reports, “Some families are so anxious to get
into public day care that they upend their lives, moving to districts known to have the shortest
waiting lists.” However, very little statistical evidence on this issue has been reported.
We provide evidence that households do not move for childcare, using information on movingfrom the Employment Structure Survey (ESS) 2002. ESS is conducted by the Statistics Bureau
every three years and covers about 1% of the population. In the 2002 survey, questions about
moving are asked in the ESS. We take a sample of mothers of children under 6 and examine theirreasons of the most recent move and where they moved from. The sample size is 41,677.
The first key question is, “Why did you move to the current place?” The respondents are asked
to choose one of 13 alternatives that includes “for childrearing and education.” The second key
question is, “Where did you live before the most recent move?” We find that only 0.8% of themothers moved from “other prefectures” for “childrearing and education,” although 7.9% of themothers moved within the same prefecture for “childrearing and education.”

14

15
2.576∗∗∗(0.950)
X
0.620
235

0.493
235

0.221∗∗∗(0.078)−0.250∗∗∗(0.077)−0.685∗(0.355)−2.469∗(1.367)(2)All
0.567∗∗∗(0.083)0.162∗∗∗(0.027)(1)All
0.686∗∗∗(0.077)−0.284(0.371)
X
X
0.984
235

−0.001(0.032)0.015(0.028)0.164(0.144)−0.235(0.544)(3)All−0.147(0.110)(5)All−0.222(0.228)0.396(0.269)0.025(0.024)−0.019(0.027)0.456∗∗∗(0.102)−0.313(0.307)(0.097)0.139(0.315)
X
X
0.992
235

0.039
(0.024)−0.034(0.026)
0.462∗∗∗(0.107)−0.500(0.331)(0.074)0.127(0.316)
X
X
0.992
235

(4)
All
0.117∗(0.064)0.615
235

0.134∗∗∗(0.017)(6)
Nuclear0.668∗∗∗(0.051)

0.771
235

0.839
(0.635)X

0.148∗∗∗(0.054)−0.150∗∗∗(0.051)−0.210(0.262)−0.435(1.064)(7)
Nuclear0.516∗∗∗(0.064)0.018(0.260)
X
X
0.994
235

−0.018(0.020)0.013(0.021)0.250∗∗∗(0.087)−0.283∗∗(0.141)(8)
Nuclear0.322∗∗∗(0.058)0.218
235

0.342∗∗∗(0.050)(9)3-Gen0.515∗∗∗(0.137)0.651
235

4.448∗∗∗(0.803)X

0.317∗∗∗(0.071)−0.366∗∗∗(0.065)−1.313∗∗∗(0.287)−3.135∗∗∗(0.951)(10)
3-Gen0.439∗∗∗(0.093)0.799∗∗(0.396)
X
X
0.977
235

0.037
(0.039)−0.032(0.039)−0.324(0.341)−1.297(1.050)(11)
3-Gen−0.243(0.158)Source: Census 1990-2010
Note: The dependent variable in all models is the maternal employment rate at the prefecture level. The number of households is used as weight. Standard errors
are clustered at the prefecture level.
Technical Note: The model is estimated using the R language and the regression tables are prepared using the software by Leifeld (2013).

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

Year FE
Prefecture FE
R2
Num. obs.

(Intercept)

Unemp Rate

Husband’s % Empl.

Husband’s Age

Age

Cap./U6-Pop. × % Nuclear HH

Household TypeCap./U6-Pop.

Table 2: Effects of Childcare Availability on Employment RateOur analysis using the ESS suggests that inter-prefecture migration for childcare is quite uncommon. Despite that, the media often reports interviews with parents who are desperate for theaccredited childcare service. This might look somewhat surprising to some readers, but we consider this to be reasonable given that the cost of an inter-prefecture move can be high. The costof inter-prefecture moving includes not only moving expenses but also the loss of social networks,
familiarity with the local public services, proximity to the current workplace, etc. Perhaps, moving
for childcare is beneficial for the highly skilled women despite the significant cost, but only 21%
of married women age 30-34 have a bachelor’s degree or higher, according to the Japanese 2010
census. We conclude that inter-prefecture migration for childcare is rare and it would not bias theestimate of the causal effect of childcare availability.
4.3.2

Effects of Other Family Policies and Labor Demand Shocks

Childcare expansion is only one of many pro-family policies, and pro-family policies other thanchildcare expansion are likely to raise the employment of mothers of children under 6. Because
childcare expansion and other pro-family policies are developed by the similar value and philosophy, changes of these policies may be correlated, which leads to an upward bias for the estimatedeffects of childcare availability. We are not aware of any major prefecture-level policies that arelikely to affect the maternal employment in general, but several small policies may collectively
affect it.19
There may also be changes in other prefecture-level labor market conditions that affect maternal employment. For example, some industries that employ many women (e.g. service industries)
grew significantly, and the growth of these industries may be correlated with the growth of childcare availability. If this is the case, the estimated effects of childcare availability are upward biased.
To address this issue, we estimate the effects of childcare availability on maternal employment
for households with children age 6-14.20 Family policies other than childcare policies and the labordemand shocks from service industries are likely to affect not only mothers of children under 6,
but also mothers of children age 6-14, because their labor market skills are similar. However, themothers of children age 6-14 are not strongly affected by childcare availability, because these older
children are not eligible for the accredited childcare service. If strong positive effects are found
for the mothers of older children, our main results above are likely to be driven not by childcare
availability, but by an omitted factor that affects all mothers regardless of the age of the youngest
child.
Table 3 shows the regression results for the employment rate of mothers with children aged 6 to
19 Nakajima and Tanaka (forthcoming) report that pronatal policies are considerably different across municipalities.
20 The employment of parents is reported by a different age-window across census years. The employment variable

can be constructed from 1990 to 2010 for 3 age-windows including 0-5, 6-14, and 15-17.

16

14. When prefecture effects are controlled along with other observed characteristics, the estimatedcoefficients are either negative or insignificant. These results indicate that maternal employment
for households with children age 6-14 does not respond to childcare availability, which in turnsupports the claim that our main result in Section 4 is indeed driven by childcare availability.
Although we believe that the two most important sources of an endogeneity bias are ruled out,
our estimates may be biased if a supply or demand shock for young mothers is correlated with thechildcare availability index. Note that this supply or demand shock must be specific to mothersof children under 6. If a supply or demand shock was correlated with childcare availability andcommon to young and old mothers, we should have detected it in the exercise in Section 4.3.2.
This endogeneity bias is possible when the local governments increase their childcare capacity
in response to the requests from mothers of children under 6 who want to participate in the labor
force. Unfortunately, we do not know the extent to which this mechanism biases our estimates.
While that is technically possible, we do not consider the possibility of a demand shock thatwas specific to mothers of children under 6 and correlated with childcare availability. Employers
are unlikely to distinguish young and old mothers, because they are very similar in skills andcommitment to work.

17

Table 3: Effects on the Employment of Mothers of Children Age 6-14
(1)All
−0.090(0.140)Age−0.000(0.024)
Husband’s Age−0.030(0.032)
Husband’s % Empl.
0.274(0.270)
Unemp Rate
−0.949∗∗(0.420)(Intercept)
1.651∗∗(0.717)R2
0.986
Num. obs.
235
Year FE
X
Prefecture FE
X
Household TypeCap./U6-Pop.

(2)(3)
Nuclear3-Gen−0.001 −0.206∗∗(0.121)(0.103)0.016−0.017(0.022)(0.019)−0.016−0.003(0.025)(0.026)0.580∗∗−0.268(0.232)(0.374)−0.303 −1.762∗∗∗(0.326)(0.632)0.048
1.847∗∗(0.620)(0.808)
0.985
0.983
235
235
X
X
X
X

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

Source: Census 1990-2005
Note: The dependent variable in all models is the maternal employment rate at the prefecture level.
The number of households is used as weight. Standard errors are clustered at the prefecture level.
4.3.3

Other Issues

Sample Selection We take the marital status and the presence of children in the family as given.
Our results are valid as long as the composition of this selected group is stable, but it may not
be invariant to the change of childcare availability. This selection is likely to be positive in thesense that more women are selected into motherhood as childcare becomes more available. We do
not know precisely the extent to which this selection affects our argument. However, we did not
find evidence for the positive selection, when we regress the fertility rate on childcare availability
and other control variables. Even if the selection is a real concern, we consider it informative to
examine the responses by those who are likely to be most strongly affected by childcare availability.
If they do not respond to childcare availability, effects on other group (e.g. single women) is likely
to be null.
Prefecture-Specific Trend Not all of our results are robust to the inclusion of a linear trend.
Indeed, only the result for nuclear households is robust. Our preferred choice is without a linear

18

trend. With only five data points, there is not much nonlinearity in the prefecture-specific growthof childcare availability (remember that we include year dummy to control for nonlinear growth at
the national level). It is likely that we pick up noises when the prefecture-specific linear trend isincluded, rather than the true relationship. The inclusion of a linear trend also makes interpretation
very complicated.
The main concern from not using a linear trend is an endogeneity bias. One might argue that
our result is driven by a trend, rather than the causal relationship that we claim. We address thisconcern by taking a control group of families of children 6-14. Our placebo or falsification test
indicates that families of children 6-14 did not respond to childcare availability. This success inthe placebo test makes our results credible, even though a linear trend is not included.

4.4

Comparison with Estimates From Other Countries

Our results are in line with those by Fitzpatrick (2010) and Havnes and Mogstad (2011) who
find that an expansion of childcare crowded out informal childcare arrangements in the U.S. and
Norway, respectively. In contrast, ? report that crowding out did not occur in Spain where someof the labor market institutions are comparable to those of Japan. For example, the gender gap islarge, the female labor force participation rate is low, and the traditional family values are rooted
in these two countries. Given these similarities, our results may seem somewhat counterintuitive.
The key difference between Japan and Spain, however, is prevalence of three-generation household. Iacovou and Skew (2010) report that, in Spain, only about 6% of children under 18 lived with
grandparents in 2007. This number is about the average of the OECD countries, which is 6.6%.
Although we cannot construct exactly the same statistic for Japan, we find that 22.6% of children
under 20 (instead of 18) live with their grandparents, according to the 2005 census. This number
is one of the highest among the OECD countries. The high share of three generation households inJapan explains why the maternal employment rate does not increase with childcare availability, despite the fact that female labor force participation rate is low. The comparison of the results acrosscountries suggests that prevalence of informal childcare arrangement is the major determinant of
the effect of childcare availability on maternal employment.

5Effects on Household Structure

An economic benefit for the parents of forming a three-generation household is the informal childcare provided by the grandparents. However, if quality childcare is provided by the local government at an affordable price, this economic benefit of a three-generation household decreases. As aresult of the improved childcare availability, some parents may want to form a nuclear household
19for more privacy and independence. Whether childcare expansion affects the household structure
or not is an important question for the welfare of families beyond maternal employment.
To explore this issue, we regress the nuclear household share on the childcare availability index.
Table 4 reports the estimation results. With none or some of the observed control variables including year dummies, the coefficient is negative (columns 1 and 2). When prefecture dummies areadded, the coefficient turns positive and significant (column 3) at 0.518. Given that the childcare
availability index grew by about 0.1 from 1990 to 2010, the share of nuclear households increased
by about 5 percentage points (0.518 * 0.1), which accounts for about a third of the growth in the
nuclear household share during the two decades.
Even though the year and prefecture fixed effects are controlled, this conditional correlation
between the share of nuclear household and childcare availability may not immediately imply thecausal relationship. Given that the increasing share of nuclear household is a secular trend acrossthe country, the co-movement of the two variables may be merely a coincidence. For example,
changes in the preference for forming a three-generation family may have affected the demand for
accredited childcare.
To address this concern, we estimate the effects on families of which youngest child is aged
between 6 and 14. If the change in the preference is the driving factor, the estimated coefficient
should be similar to that for families of children under 6. The result in column 4 indicates thatthe estimate is statistically insignificant at 0.109. Given that the positive significant correlation
is found for only families of children under 6, the evidence supports our claim that childcare
availability increased the share of nuclear household. Our result is not robust to the inclusion ofprefecture-specific linear trends, but, as discussed in 4.3.3, the regression with prefecture-specifictrend is unlikely to be informative given that there are only five data points per prefecture. We try
to establish the causal relationship by conducting the placebo test in this subsection.

20

Table 4: Effects of Childcare Availability on Nuclear Household Share(1)
Age of the Youngest Child
0-5Household TypeAll
Cap./U6-Pop.
−0.341∗∗∗(0.128)Age
Husband’s Age

(2)
0-5
All
−0.327∗∗(0.128)−0.060(0.112)0.151(0.110)

(3)
0-5
All
0.518∗∗∗(0.132)0.080∗(0.046)−0.096∗∗(0.042)

(4)
6-14
All
0.109
(0.132)0.023(0.046)−0.018(0.042)(0.521)
5.077∗∗∗(1.464)−3.161∗∗(1.598)X(0.291)−0.521(0.992)0.807(0.825)
X
X
0.983
235

(0.291)
0.392(0.992)0.081(0.825)
X
X
0.992
235

Husband’s
Unemp Rate
(Intercept)
Year FE
Prefecture FE
R2
Num. obs.

0.886∗∗∗(0.044)

0.069
235

0.560
235

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

Source: Census 1990-2010
Note: The dependent variable is the nuclear household share among all households with children
under 6. The number of household is used as weight. Standard errors are clustered at the prefecturelevel.

6

Changes in Effects of Childcare Availability

We examine if effects of childcare availability on maternal employment change over time. Specifically, we estimate the fixed effect model for the periods 1990-2000 and 2000-2010 separately
and compare the results for the two decades. This comparison may provide an insight useful for
predicting effects of a further expansion of accredited childcare in the future.
Table 5 shows the results for the period 1990-2000, and Table 6 shows the results for theperiod 2000-2010. The overall effect on maternal employment is negative, but insignificant in the
1990’s, while it is positive significant at the 10% level in the 2000’s. Our analysis in the previous
section shows that the substitution of accredited childcare for informal care by grandparents offsets
positive effects of childcare on maternal employment. To see how the substitution pattern haschanged, we estimated the effects of childcare availability on the nuclear household share. The
estimated coefficient for childcare availability is 0.535 in the 1990’s, but it decreased to about21

a half at 0.271 in the 2000’s. This indicates that the substitution effect is weakening over time,
presumably because the share of nuclear household approaches to the possible highest level.
We also estimate the effects by household type. The estimated coefficients for nuclear households are very similar in two decades: 0.284 in the 1990’s and 0.295 in the 2000’s. In contrast,
there is a notable change in the effect on three-generation families. The estimate increases from
-0.272 in 1990’s to 0.090 in 2000’s. One possible explanation for this rise is that more and moregrandparents work in the labor market and do not take care of children even when they live in thesame household.
The comparison of the results between 1990’s and 2000’s indicates that the effects childcare
availability on maternal employment increased for two reasons. First, the substitution of accredited
childcare for informal care by grandparents is weakening. This is presumably because the nuclear
household share is already very high. Second, the effects on three-generation households are
increasing. This may be because recent grandparents in the 2000’s work in the labor market rather
than taking care of the children. If this trend continues, a further expansion of accredited childcare
may increase maternal employment in the future.
Table 5: Effects of Childcare Availability in 1990s
(1)

(2)
(3)
Maternal Emp. Rate
All
Nuclear3-Gen∗∗−0.2680.284−0.272(0.216)(0.122)(0.293)∗∗Age−0.027 −0.050
0.032(0.030)(0.020)(0.032)
Husband’s Age
0.015
0.036∗−0.037(0.035)(0.019)(0.044)
Husband’s % Empl. −2.072∗∗
1.134−2.547∗∗∗(0.881)(0.696)(0.711)
Unemp Rate
−0.895−0.042−1.508(0.859)(0.440)(0.932)∗∗(Intercept)
2.795−0.6433.327∗∗∗(1.330)(0.739)(0.975)
Year FE
X
X
X
Prefecture FE
X
X
X
R2
0.990
0.995
0.990
Num. obs.
141
141
141
Dependent Variable
Household TypeCap./U6-Pop.

(4)% Nuclear HH
All
0.535∗∗(0.218)0.048∗(0.028)−0.087∗∗(0.033)5.307∗∗∗(1.032)−0.087(0.843)−3.230∗∗(1.447)
X
X
0.995
141

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

Source: Census 1990-2000
Note: The dependent variable in all models is the maternal employment rate at the prefecture level.
The number of households is used as weight. Standard errors are clustered at the prefecture level.
22

Table 6: Effects of Childcare Availability in 2000s
(1)
(2)
(3)
Maternal Emp. Rate
All
Nuclear3-Gen
0.124∗ 0.295∗∗∗
0.090(0.075) (0.054) (0.114)
Age−0.0070.023−0.014(0.038) (0.038) (0.036)
Husband’s Age
0.025−0.0170.038(0.036) (0.037) (0.032)
Husband’s % Empl.
0.129
0.136
0.159
(0.091) (0.085) (0.114)
Unemp Rate
−0.473∗ −0.160 −0.669(0.277) (0.200) (0.603)(Intercept)−0.461 −0.087 −0.660∗(0.372) (0.215) (0.333)
Year FE
X
X
X
Prefecture FE
X
X
X
2
R
0.996
0.997
0.994
Num. obs.
141
141
141
Dependent Variable
Household TypeCap./U6-Pop.

(4)% Nuclear HH
All
0.271∗∗(0.121)0.120∗∗(0.054)−0.112∗∗(0.052)0.261∗∗(0.122)0.661(0.436)0.489(0.623)
X
X
0.994
141

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

Source: Census 2000-2010
Note: The dependent variable in all models is the maternal employment rate at the prefecture level.
The number of households is used as weight. Standard errors are clustered at the prefecture level.

7

Conclusion

Using prefecture panel data from the Japanese quinquennial census from 1990 to 2010, we estimate the effects of childcare availability on maternal employment. We depart from previous
papers on Japan by controlling for prefecture fixed effects and find that the estimates identified
by cross-sectional variations alone can be severely biased. Indeed, contrary to popular belief,childcare availability is uncorrelated with maternal employment when prefecture fixed effects arecontrolled. Evidence suggests that this is because households substitute the accredited childcare
service for informal childcare by grandparents, as more and more households choose not to livewith grandparents.
We also analyze nuclear and three-generation households separately. The estimation resultshows that the employment rate of mothers from nuclear households increases with childcare
availability, whereas that of mothers from three-generation households does not. We address two
potential sources of a simultaneity bias and find them negligible. First, we show evidence that inter23

prefecture migration for childcare is rare. Second, we show that the employment rate of mothersof children 6-14 does not change with childcare availability. These results implies that prefecturelevel policy changes and other shocks common to young and old mothers are not correlated withthe changes of childcare availability.
Can further expansions of childcare raise the maternal employment rate? The answer to thisquestion is not immediately obvious, but our analysis in Section 6 shows some promise. The key
reason for the null effect is the substitution of accredited childcare for informal care by grandparents, but this crowding-out effect is weakening over time. We estimate effects of childcare
availability separately for the 1990’s and the 2000’s, and find that the employment effect became
positive in the 2000’s, although it is modest. Given that the share of three-generation households
was already at a low 13.5% in 2010, the overall effect of childcare availability on maternal employment may be getting stronger. Moreover, even three-generation families may positively respond to
the expansion of childcare, because recent grandparents may work in the labor market rather thanto take care of children.
A catch is that the pace of childcare expansion has to exceed the child population growth inorder to keep childcare availability improving. More than 70% of the growth of childcare availability in 1990-2010 is explained by the decreasing child population. If the fertility rate starts toincrease due to pro-family policies and other factors, childcare availability will not improve unlessthe pace of expansion is fast enough. Even larger government expenditure than the current one
may be necessary to expand the supply of accredited childcare service.
The main limitation of the current paper is that we cannot examine detailed labor market outcomes such as wage, earnings, hours of work, and job type, due to lack of data. Understanding
the effects of childcare availability on these outcomes will shed light on welfare consequences andopen the door to a cost-benefit analysis. We leave these interesting issues for future research.

References
A BE , Y. (2013): “Regional variations in labor force behavior of women in Japan,” Japan and theWorld Economy, 28(0), 112 – 124.
A SAI , Y. (2015): “Parental leave reforms and the employment of new mothers: Quasiexperimental evidence from Japan,” Labour Economics, (0), –.
A SAI , Y., R. K AMBAYASHI , AND S. YAMAGUCHI (2014): “Childcare Availability, Household
Structure, and Maternal Employment,” Department of Economics Working Papers 2014-08, McMaster University.

24

BAKER , M., J. G RUBER , AND K. M ILLIGAN (2008): “Universal Child Care, Maternal Labor
Supply, and Family Well-Being,” Journal of Political Economy, 116(4), 709–745.
B ERLINSKI , S., AND S. G ALIANI (2007): “The effect of a large expansion of pre-primary school
facilities on preschool attendance and maternal employment,” Labour Economics, 14(3), 665 –680.
B RILLI , Y., D. D EL B OCA , AND C. P RONZATO (2013): “Does child care availability play a role inmaternal employment and children’s development? Evidence from Italy,” Review of Economicsof the Household, pp. 1–25.
F ITZPATRICK , M. D. (2010): “Preschoolers Enrolled and Mothers at Work? The Effects of Universal Prekindergarten,” Journal of Labor Economics, 28(1), 51–85.
G ELBACH , J. B. (2002): “Public Schooling for Young Children and Maternal Labor Supply,”
American Economic Review, 92(1), 307–322.
G OUX , D., AND E. M AURIN (2010): “Public school availability for two-year olds and mothers’
labour supply,” Labour Economics, 17(6), 951–962.
H AVNES , T., AND M. M OGSTAD (2011): “Money for nothing? Universal child care and maternal
employment,” Journal of Public Economics, 95(11), 1455–1465.
H IGUCHI , Y., T. M ATSUURA , AND K. S ATO (2007): “Effects of Regional Factors on Fertility
and Employment Continuity of Wives,” RIETI Discussoin Paper No. 07-J-012 (in Japanese).
I ACOVOU , M., AND A. S KEW (2010): “Household Structure in the EU,” Institute of Social andEconomic Research, Colchester, Working Paper, No. 2010-10.
L EE , G. H., AND S. P. L EE (2014): “Childcare availability, fertility and female labor force participation in Japan,” Journal of the Japanese and International Economies, 32(0), 71 – 85.
L EFEBVRE , P., AND P. M ERRIGAN (2008): “Child-Care Policy and the Labor Supply of Motherswith Young Children: A Natural Experiment from Canada,” Journal of Labor Economics, 26(3),519–548.
L EIFELD , P. (2013): “texreg: Conversion of Statistical Model Output in R to LATEX and HTML
Tables,” Journal of Statistical Software, 55(8), 1–24.
M INISTRY OF H EALTH , L., AND W EALTH (2009): “Summary of the Survey of Local Child
Welfare Services,” available from http://www.mhlw.go.jp/toukei/saikin/hw/jidou/09/dl/80.pdf.

25

NAKAJIMA , R., AND R. TANAKA (forthcoming): “Estimating the Effects of Pronatal Policies on
Residential Choice and Fertility,” Journal of The Japanese and International Economies.
N IHON -J IDOU -M ONDAI -C HOUSAKAI (1978): Jidou Fukushi Sanjuu Nennno Ayumi (30 Years ofChild Welfare). Division of Children and Family, the Ministry of Welfare.
N OLLENBERGER , N., AND N. RODRIGUEZ -P LANAS (2015): “Full-time universal childcare in acontext of low maternal employment: Quasi-experimental evidence from Spain,” Labour Economics, (0), –.
S ASAKI , M. (2002): “The Causal Effect of Family Structure on Labor Force Participation among
Japanese Married Women,” Journal of Human Resources, 37(2), 429–440.
S HIGENO , Y., AND Y. O OKUSA (1999): “The Effects of Childcare Policy on Fertility and Employment,” Kikan Shakai Hosho Kenkyu, 35, 192–207 (in Japanese).
TABUCHI , H. (2013): “Desperate Hunt for Day Care in Japan,” New York Times, Online; PublishedFebruary 26, 2013.
U NAYAMA , T. (2011): “Compatibility of Marriage, Fertility, and Work and Childcare Availability,” Nihon Keizai Kenkyu, 65, 1–22 (in Japanese).
YAMAGUCHI , S. (2014): “Dynamic Effects of Parental Leave Policy on Female Employment,”
McMaster University.
YAMAKAWA , R. (2007): From Security to Mobility? Changing Aspects of Japanese Dismissal
LawUniversity of Washignton Press, pp. 483–520.
YOKOHAMA (2014): “The Number of Waitlisted Children for Childcare As of April 1, 2014,”
Press Release.

A

Additional Table

We show evidence that whether the actual or potential capacity rate is used does not change ourresults. The actual capacity rate is defined as the ratio of capacity to population of children under
6 and used in the body of the paper. The potential capacity rate is proposed by Unayama (2011)
and defined as the ratio of capacity to population of women aged between 20 to 44. Table 7 iscomparable to Table 2, while Table 8 is comparable to Table 4. The coefficients and standard
errors become about three times greater when the potential capacity rate is used, but they are very
similar otherwise.
26

We also regress the actual and potential capacity rates on the year and prefecture dummies.
The correlation coefficient of the two vectors of residuals is 0.92, which accounts for why results
are not very different.

27

28
0.171∗∗∗(0.020)
0.578
235

(1)
All
1.912∗∗∗(0.185)1.923∗(1.102)
0.674
235
X

0.195∗∗∗(0.069)−0.204∗∗∗(0.071)−0.789∗∗(0.349)−2.139∗(1.090)(2)All1.644∗∗∗(0.237)−0.237(0.379)
0.984
235
X
X

0.003
(0.032)0.010(0.029)0.146(0.149)−0.275(0.550)(3)All−0.379(0.349)(5)All−0.193(0.443)0.232(0.166)0.034(0.023)−0.029(0.026)0.454∗∗∗(0.104)−0.417(0.289)(0.091)
0.165(0.326)
0.992
235
X
X

0.035
(0.024)−0.029(0.027)0.464∗∗∗(0.109)−0.472(0.327)(0.071)
0.077(0.322)
0.992
235
X
X

(4)
All
0.327
(0.208)0.158∗∗∗(0.015)
0.609
235

(6)
Nuclear1.714∗∗∗(0.145)0.285(0.784)
0.809
235
X

0.121∗∗∗(0.044)−0.107∗∗(0.045)−0.300(0.252)−0.272(0.789)(7)
Nuclear1.495∗∗∗(0.194)−0.064(0.270)
0.994
235
X
X

−0.028(0.021)0.024(0.021)0.269∗∗∗(0.093)−0.233(0.177)(8)
Nuclear0.915∗∗∗(0.214)0.323∗∗∗(0.043)
0.314
235

(9)
3-Gen
1.617∗∗∗(0.336)

4.168∗∗∗(0.865)
0.657
235
X

0.297∗∗∗(0.073)−0.342∗∗∗(0.068)−1.268∗∗∗(0.324)−2.888∗∗∗(0.919)(10)3-Gen
1.170∗∗∗(0.245)0.864∗∗(0.429)
0.977
235
X
X

0.040
(0.041)−0.036(0.041)−0.357(0.367)−1.376(1.047)(11)
3-Gen−0.637(0.482)

Source: Census 1990-2010
Note: The potential childcare availability is measured by the ratio of capacity to the population of women aged between 20 and 44. The dependent variable in allmodels is the maternal employment rate at the prefecture level. The number of households is used as weight. Standard errors are clustered at the prefecture level.

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

R2
Num. obs.
Year FE
Prefecture FE(Intercept)

Unemp Rate

Husband’s % Empl.

Husband’s Age

Age

Cap./Female-Pop. × % Nuclear HH

Household TypeCap./Female-Pop.

Table 7: Effects of Potential Childcare Availability on Employment Rate

Table 8: Effects of Potential Childcare Availability on Nuclear Household Share(1)
Cap./Female-Pop. −1.438∗∗∗(0.311)
Age
Husband’s Age(2)−1.111∗∗∗(0.324)−0.042(0.106)0.116(0.105)(3)
1.412∗∗∗(0.459)0.064(0.047)−0.078∗(0.044)(0.494)4.728∗∗∗(1.292)−2.544∗(1.537)
X(0.291)−0.395(1.071)0.627(0.857)
X
X
0.983
235

Husband’s
Unemp Rate
(Intercept)

0.929∗∗∗(0.038)Year FE
Prefecture FE
R2
Num. obs.

0.186
235

0.592
235

∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1

Source: Census 1990-2010
Note: The potential childcare availability is measured by the ratio of capacity to the population of
women aged between 20 and 44. The dependent variable is the nuclear household share among all
households with children under 6. The number of household is used as weight. Standard errors are
clustered at the prefecture level.

29