Abstract
I exploit the introduction of a policy package in France aimed at helping parents with the care of young children. The reform affected all households with pre-school age children and had two dimensions: a short stay-home subsidy for first-time mothers wishing to take-up parental leave and an increase in childcare subsidies for parents using childminders—the main formal care option in France. Importantly, policymakers did not explicitly intervene in the childcare infrastructures. I rely on a diff-in-diff empirical strategy to evaluate the labour market outcomes of mothers with pre-school age children in the short-run and the long-run. The reform had negligible effects in the short-run. In the long-run though, first-time mothers—and particularly the lower-educated group—took advantage of the parental leave subsidies to reduce their employment rate. This freed up formal childcare places and allowed middle-class educated mothers of two children to use the more generous childcare subsidies and therefore work more. The fact that the effects take time to materialise and do not appear at the aggregate level for the targeted population suggests that the policy did not induce any net increase in the supply of care places and simply led to a re-allocation of care modes among mothers of pre-school age children.
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Notes
This was the main type of formal childcare in France at the time of the reform.
I also check that father’s labour supply and fertility choices were not impacted by the reform.
Prior to the reform this benefit system was called APE (“Allocation Parentale d’Education”), with the reform the new benefit system was given the name of PAJE (“Prestation d’Accueil du Jeune Enfant”).
The maternity leave is normally 6 weeks pre-birth and 10 weeks post-birth, but 3 weeks pre-birth can be substituted for 3 weeks post-birth, bringing the maximum duration to 13 weeks post-birth.
An assistante maternelle is “accreditated” (allowed) to look after a certain number of children by the local authorities. The number of children to care for increases with seniority and other criteriae.
They are more expensive than childminders and the subsidies for this type of care were affected by the reform.
Such information is only available for the years 2002 and 2007 around the reform and come from the household surveys described in the last section of the article.
Having worked for two years prior to the child’s birth, within a timeframe that depends on the number of existing children in the family.
These childcare subsidies are also available for children between the age of three to six but are about half the value of those for children below three years old. Most of these children go to school, so these subsidies are less popular. I have looked at this group, but did not find any significant impact of the subsidies’ changes and decided to focus only on pre-school children.
Note that the government also pays 100% of the employer social contributions.
To be eligible, one needed a monthly salary of at least €374 if a lone parent and €748 if in a couple.
Their dataset did not allow them to study working hours, wages or disentangle reported spending on official childcare between hourly cost and hours used of formal care. This last point should be important as the aim of subsidising childcare is generally to achieve cheaper hourly cost for parents or a higher number of hours used. Simply increasing the revenue of childcare providers via subsidies to parents is not usually the aim of such policies.
Prior to 2003, the surveys were collected once a year and were called Enquête Emploi. In 2003, it was replaced by a quarterly rolling panel (Enquête Emploi en Continu) where each household was interviewed for six consecutive quarters. The procedure for data collection was very similar between the two surveys and I control for potential differences in my specification.
Because mothers might combine the maternity leave with some normal vacation leave, I define the group cut-off as having a child up to 11 months old instead of nine and half months old (13 weeks of maternity leave plus 26 weeks of stay-home benefits).
Even though children can and usually go to school from the age of 3 onwards, households may still employ private carers to look after children outside of school hours for those younger than 6 when childcare subsidies are still available to parents.
This is particularly necessary in the short-run period as the child age composition of the treated group evolves throughout the quarters and years. For instance, in the year 2004, only the mothers with a child younger than one are receiving the new benefit and are hence in the treatment group.
For brevity, I do not report graphs for the other groups and variables as they confirm the absence of a significant impact from the policy. The graphs are available upon request and can be seen in a preliminary draft of this paper in chapter 1 of de Muizon (2014).
I use the local administrative area called “departement” that can usually be crossed within a couple of hours of driving. I did not include this control in the main specification because this is an imperfect proxy as it doesn’t fully match with households commuting area if they live close to departmental borders.
The checks presented here cannot be performed for the long-run period. By 2007, all children below the age of three would be born after 1 January 2004, implying that their mothers would automatically be under the new benefit regime. In chapter 2 of Jourdain de Muizon (2014) I used a structural labour supply model to simulate the impact of the stay-home subsidy on first-time married mothers’ work choices. The results suggested that the long-run impact of the policy would reduce their employment rates by slightly more than 11% points, a similar order of magnitude to the one found here in Table 8. Also, the structural model approach predicted very little effect on working hours, in line with the results presented in the main specification.
For instance, in July 2005 a mother whose youngest is 1 to 3 years old but born before 2004 would be in the control group, while the mother of a child born in 2004 would be in the treatment group.
I run a regression similar to Persson and Rossin-Slater (2019), that takes the following form:
$${y}_{itp}=\alpha +{\beta }_{1}1[t\ge c]+{\beta }_{2}{R}_{i}* 1[t\ge c]+f(t-c)+1[t\ge c]* f(c-t)+{\theta }_{p}+\beta ^{\prime} {X}_{i}+{e}_{itp},$$where yitp represents the mother’s labour market outcome, Ri is an indicator set to 1 for children in the reform sample (i.e., born 2 quarters on each side of 1 January 2004), c denotes January 1 of any observation period, the dummy variable 1[t ≥ c] is set to one for children born in the first two quarters of the year, f() is a linear function controlling for different trends in the period before and after the policy cut-off date. I include fixed effects for each of the three periods of observation θp, note that the main effect of Ri is therefore absorbed by the period dummy. The months before January 1 are represented as (c − t),and after as (t − c).The control variables Xi are the set of controls used in the main specification. The coefficient of interest capturing the impact of the policy is β2. I restrict the analysis to women with a child born within a 2 quarters window around the policy cut-off date.
Looking at confidence intervals and using simple Z test.
In these two robustness checks, the treated mothers are compared to a group of mothers whose youngest child is in the same age bracket but were not eligible to the new policy. This is a similar approach to Joseph et al. (2013) or Givord and Marbot (2015). As discussed in the literature review, their approach may bias the estimates if the non-treated mothers were indirectly affected by the policy changes. If the non-treated mothers found it harder to find childcare places for example, this may negatively affect their labour supply and could result in an apparent positive effect of the reform on treated mothers’ labour supply. But this estimated positive effect would disappear when the treated mothers are compared to a group of mothers not impacted by the reform in any way, as in my baseline specification.
“Enquêtes mode de garde et d’accueil des jeunes enfants” from the DREES (Directorate of Research, Study, Evaluation and Statistics) agency.
I focus here on mothers whose youngest is between 1 and 3 years old as they were the most affected by the childcare subsidy changes in the main analysis. To check the arguments made in that paragraph and extend the analysis to mothers whose youngest was below one, I estimated probabilities of working and using a specific childcare mode by following a strategy similar to Baker et al. (2008). The results are presented in the appendix.
To confirm these findings, I study the evolution of childminders’ wages in the Labour Force Surveys. The hourly wage of childminders moved from 3.27 € in 2002 to 4.50 € in 2007 and 5.48 € in 2009, which represents a 24% increase in real terms by 2007 and 46% by 2009 when the minimum wage in real terms was increased by 11% and 13% respectively (from 6.75 € to 8.36 € and 8.77 €). Note that technical reports from French statistical institutes and agencies also found some evidence of these effects. For instance, see Marical (2007),“Les determinants des salaires des assistantes maternelles et les effets de la PAJE,” Recherches et Previsions, 88: 35–52.
This is described in more details in the yearly bulletin of the DREES, see Vanovermeir (2011), “Les prestations familiales et de logement en 2009 Les bénéficiaires des aides á la garde d’enfants plus nombreux,” Etudes et resultats, 769, DREES.
The DREES (Directorate of Research, Study, Evaluation and Statistics) agency from the French Labour Ministry reports every year the number of childcare providers at the national and local level.
Note that in 2004, the treated group is only composed of mothers whose youngest child is younger than one, explaining the sharp drop observed in that year in Fig. 11.
It should be noted that Lequien (2012) found long-term effects of the 1994 APE reform on mothers of two children’s wages. The control group in the years 2001-03 would be composed of mothers who were not affected by the APE (those whose youngest child was born before 1994 and who would be 8 or 9 years old in 2001) and of mothers who were affected (those whose child was born in 1994 and later). After 2003, the control group is exclusively composed of mothers affected by the APE policy. Evidence presented here using the Labour Force Surveys point to limited risks of such a bias. Also, observable characteristics such as age, marital status, education and labour supply across the control group did not differ much whether mothers would have been affected nearly a decade earlier by the APE or not.
In 1985, the French government set the objective that 80% of each 18 years old cohort would graduate from high-school by 2000. The objective was eventually reached in 2012.
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Acknowledgements
I am particularly grateful to Guy Laroque, Richard Blundell, Magne Mogstad, Ian Walker, Ian Preston, Ana Raute, Thibaut Lamadon, Yves Gueron, Yehui Wu, two anonymous referees and the editors of the journal for their constructive comments and suggestions. All errors and omissions remain my own.
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Appendix
Appendix
1.1 Details of the changes to the parental stay-home subsidies
1.2 Benefit take-up
The Agency in charge of allocating and paying the family-related benefits CNAF (“Caisse Nationale d’Allocations Familiales”) reports every year the number of claimants to each of its benefits. This data is the source used to construct the graphs presented here. Prior to the reform, the female labour supply was generally increasing in France, and as a result, the number of claimants to childcare subsidies for private carers was increasing as shown in Fig. 8. Figure 9 reports the number of claimants in France to the stay-home subsidies for all types of households. The increase in the number of claimants in 2004 and 2005 by about 20,000 for the fully out-of-work transfer is mainly attributable to first-time mothers becoming eligible to the benefit (see Joseph et al. 2013). Each year there are about 300,000 first-time mothers in France. The generosity of that benefit was not modified for mothers of two and more children (Table 5 above). For mothers choosing to reduce their working hours post-birth and claim the part-time stay-home subsidy, about 10,000 first-time mothers claimed it and the large increase observed in Fig. 9 is driven by mothers of two or more children. Surprisingly, the number of mothers claiming the benefit to completely stay out of work started to fall after 2006. This is due to a fall in the number of mothers with more than one child choosing to stay home fully while the number of first-time mothers claiming the full subsidy remained broadly constant.Footnote 28
Number of childcare subsidies claimants
Number of claimants to the stay-home subsidies
1.3 Summary statistics and regression results table
1.4 No reform-induced rise in childminders
At the aggregate level, the reform did not induce any rise in the supply of care places beyond the long-term trend associated with secular rise in female employment.Footnote 29 While the number of childminders kept rising after the policy was implemented, the growth rate was similar to the years prior to the policy change (see Fig. 10). A slight modification in the data methodology between 2001 and 2002 explains the apparent stagnation in the number of carers around that time.
Number of childminders (assistantes maternelles)
1.5 Common trend assumptions
I reproduce below time-series of the dependent variables for the treatment group (mothers with one or two children younger than 3 years old) and control group (mothers with one or two children aged 6 to 10 years old). To further check the validity of the control group and ensure that the pre-reform trends were similar between mothers of pre-school children and mothers of older children, I perform an event type of study. I split my sample between observations in the treatment group and control group. I regress the dependent variable on yearly dummies, a dummy if in the treatment group and yearly indicators interacted with the treatment dummy. I do not perform the analysis for years before 2000 as the labour market outcomes of the treatment group may still have been affected by the APE reform changes in 1994 (Piketty (2005) was identifying effects until 1998).
Along the extensive margin of work decision (Fig. 11), the common trend assumption in the decade around the reform seems to hold.Footnote 30 Figure 12 presents the coefficients and the 95% confidence interval of the yearly indicators interacted with the treatment dummy for the extensive margin of work. The yearly indicators for the treatment group are not significantly different from zero. Post 2004 though, the coefficients are concentrated below zero. This is a reflection of the large negative impact of the policy on the subgroup of first-time mothers with a child younger than one.
Common trend for treatment and control groups, employment rate
Treated vs control group trend analysis, employment rate
Along the intensive margin of work decision (Fig. 13), the common trend assumption in the years prior to the reform seems to generally hold. Figure 14 presents the coefficients and the 95% confidence interval of the yearly indicators interacted with the treatment dummy for the intensive margin of work. The yearly indicators for the treatment group are not significantly different from zero.
Common trend for treatment and control groups, working hours
Treated vs control group trend analysis, working hours
Regarding hourly wages (Fig. 15), the common trend assumption in the years prior to the reform seems to broadly hold well. Figure 16 presents the coefficients and the 95% confidence interval of the yearly indicators interacted with the treatment dummy for the logarithm of hourly wages. The yearly indicator for the treatment group are not significantly different from zero.Footnote 31
Common trend for treatment and control groups, wages
Treated vs control group trend analysis, wages
1.6 Check on demographic characteristic trends: education, household composition and age
In the graphs below I check the share of mothers that at least graduated from high school (the higher educated group defined in the main text). The common trend on that covariate does not seem to hold perfectly, especially in the last year of observation. Note that after 2005, the educational attainment of the control group rises faster than that of the treated group. There is a likely cohort effect for mothers in the control group between those observed in the early 2000’s that could have graduated from high school in the first half of the 1990’s and those observed in the late 2000’s that could have graduated from high school in the late 1990’s. In this period, public policy was explicitly aiming at gradually increasing the educational attainment in the population (see Verdugo (2014)Footnote 32) (Figs 17 and 18).
Common trend for treatment and control groups, education
Treated vs control group trend analysis, education
In the two figures below (Figs 19 and 20), I check that the reform had no impact on household formation and the share of mothers in a couple appears unaffected throughout the years of observation.
Common trend for treatment and control groups, mothers in couple
Treated vs control group trend analysis, mothers in couple
In the two figures below (Figs 21 and 22), I check that the reform had no impact on the age of the mothers. The average age of mothers in treatment and control groups seem to follow the same evolution.
Common trend for treatment and control groups, age of mothers
Treated vs control group trend analysis, age of mothers
1.7 Probabilities of working and using a childcare mode
I use the childcare surveys from the last section of the main text and run a simple difference regression (unfortunately no obvious control group exists as childcare choices mainly affect mothers of pre-school children). For each household category, I run the following regression:
where Childcareij, represents the main childcare mode j used by household i. Post is a dummy equal to 1 if the observation was after the policy change, and Xi is a vector of controls (education, age of the mother and its square, the age in months of the youngest child, as well as a dummy if married and in couple). I look at three main modes of care: private carers (childminders and nannies) that was affected by the reform, kindergarten (creches) and other. The estimates of the γ parameter above are presented in Table 12.
From this exercise, we can conclude the following:
-
(1)
A significant fall in the probability of using kindergarten for mothers of one child and an increase for mothers of two children, independent of their work situation.
-
(2)
A significant fall in the probability of using private carers while not working induced by the modification in eligibility rules to subsidies (now available exclusively to working mothers).
-
(3)
For mothers with a child aged 1 to 3 years old (irrespective of the number of children), the probability of working and using other modes significantly rose. The probability of working and using private carers rose (although the impact is not statistically significant and bigger for mothers of two children).
1.8 Labour supply of fathers
In the figures below (Figs 23–26), I report the employment rate of fathers observed in the Labour Force Surveys by the year of birth of the youngest child in the household. The graphs highlight that there was no clear impact of the reform.
Similar graphs representing the fathers’ intensive margin of work show no reaction to the reform and are not presented here. They are available upon request.
Employment rates for fathers of one child (youngest under 1 year old)
Employment rates for fathers of one child (youngest 1 to 3 years old)
Employment rates for fathers of two children (youngest under 1 year old)
Employment rates for fathers of two children (1 to 3 years old)
1.9 Fertility trends
Piketty (2005) estimates a small impact of a much longer parental policy change on fertility in France in the mid to late 1990’s. The data available does not allow me to study the response of natality choices to the policy with precision. I rely on the population statistics from the INSEE (National Institute of Statistics and Economic Studies) to construct the indicators reported in this section. The number of children born per 100 women aged 15 to 49 has been constantly rising between 1998 and 2010 in France (as shown in Fig. 27). This is particularly true for the number of first births per woman (Fig. 28). The parental leave subsidies that now became available to this group of women do not appear to have had any impact on their decision to start a family. The difference with the results in Piketty (2005) could be due to the much shorter duration of the program (6 months here versus 3 years in Piketty (2005)). Also, Joseph et al. (2013) and Givord and Marbot (2015) provided similar checks that the 2004 policy change did not impact fertility in the short-run at least.
Number of births per 100 women aged 15 to 49
Number of first births per 100 women aged 15 to 49
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de Muizon, M.J. Subsidies for parental leave and formal childcare: be careful what you wish for. Rev Econ Household 18, 735–772 (2020). https://doi.org/10.1007/s11150-020-09489-9
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DOI: https://doi.org/10.1007/s11150-020-09489-9