Childcare, Parental Behavior and Well-Being during Childhood

1 Introduction

In the last few decades, there has been a sharp increase in the labor market employment rate of mothers in developed countries. Although this has had a positive effect on family incomes, it has also made parenting a more demanding and stressful task given the increased time and pressure from work. Concurrently, parents growing demand for childcare has attracted the attention of policy makers toward public or subsidized childcare programs. The idea that childcare subsidies should no longer be limited to low-income families, but should instead be universal, as is the case in most European countries, is slowly emerging as a model for North American governments, particularly in Canada.

Studies estimating the effects of universal childcare policies have focused mainly on their impact on maternal employment and child development (see Baker 2011, for a review). However, as explained by Herbst and Tekin (2014), a full evaluation of childcare subsidies requires a thorough understanding of the ways in which subsidies influence both parents and their children. When mothers reallocate their time from home to the labor market, a change in the nature of time spent with the child occurs, affecting not only children’s well-being but also parents’ well-being and behaviors with their children. An extensive literature shows how maternal well-being affects by itself a child’s well-being and development (Currie and Almond 2011; NICHD 1999, 2003).

The first major study to examine the effects of a universal childcare policy on both children and parental well-being was conducted by Baker, Gruber, and Milligan (2008). Their study exploits a major childcare reform which took place in Canada in the late 1990s. In 1997, the government of the province of Quebec initiated the gradual implementation of a low-fee childcare policy. Childcare spaces were provided at a single low-fee of CA$5 per child per day, approximately US$3.50 at that time (CA$7 as of 2004). The reform was phased in to include all of Quebec’s children less than 6 years of age (not in publicly provided kindergarten) as of September 2000. This policy had the effect of moving a large proportion of children from informal care and maternal care to regulated childcare (Lefebvre and Merrigan 2008). For example, the number of regulated childcare spaces in Quebec increased from 78,864 in September of 1997 to 258,366 as of March, 2013 (Ministre de la Famille et des Ainés 2013). More importantly, as a result of the policy, the labor force participation of mothers increased by 7.7 percentage points in Quebec by 2003 (Baker, Gruber, and Milligan 2008). These authors also show that the reform had a negative effect on several parental and child outcomes during the preschool years. In particular, the policy had a negative effect on parents’ health and behaviors.^[1] However, these results were obtained strictly with children aged 0–4 years old and their parents. At the time of these studies only a few post-policy years of data were available. The results thus reflect the short-term impact of the reform up to 2003. It is difficult to identify the mechanisms that explain their results, but several can be considered, as discussed in the next paragraphs.

In this study, we extend the research of Baker, Gruber, and Milligan (2008) on universal childcare and parental outcomes in two ways. First, we estimate the effects during the preschool years over a longer observation period using six years of additional data, up to 2009. Second, we examine whether the effects on parents found in the preschool years persist when the child enters school.

Documenting the long-term effects of universal childcare on parents is crucial to our understanding of the overall impact of such reforms, both when the child is in childcare but also once the parents are no longer directly affected by the program. Studies of long-term impacts of early childhood education and care (ECEC) on parents are very scarce and mainly focus on low-income or single-mother households (Barnett 2008; Herbst and Tekin 2014; Martin et al. 2008).^[2] Regarding Quebec’s childcare policy, the long-term effects documented focused on labor supply (Lefebvre, Merrigan, and Verstraete 2009) and child outcomes (Baker, Gruber, and Milligan 2019; Haeck, Lebihan, and Merrigan 2018),^[3] but not on parents and family interactions.

Relative to the first objective of this study, our empirical strategy allows us to measure whether the effects on the parents of preschool children found in previous research are transitional or persist when the program matures. We use data on parents of children who were eligible for low-fee childcare for up to 5 years. In previous studies, accessible data was mainly focused on the early years of the program, when children had been eligible for low-fee childcare for only 1–2 years. At this time, the number of spaces was highly constrained (Haeck, Lefebvre, and Merrigan 2015). As a result, the estimated short-term effects in Baker, Gruber, and Milligan (2008) may be different from the longer term effects once the number of spaces has stabilized and children attend childcare for 4–5 years. During the transition, parents also had to adjust to a new social norm. This may also have taken some time and may have influenced the early measured impacts of the childcare reform. Haeck, Lefebvre, and Merrigan (2015) show that the positive effects on maternal labor force participation start in 2000 at 5 percentage points, increase to 13 percentage points by 2004 and persist up to 2009, the final year^[4] of the data set. This further suggests that long term effects on parental well-being might differ.

Relative to the second objective, there are several reasons to assume that the childcare policy could affect parents with children even when they are no longer in childcare. First, Lefebvre, Merrigan, and Verstraete (2009) show that the policy had long-term labor force participation effects (when their child is in school) on mothers who benefited from the program when their child was eligible at an early age. They show that the policy had a positive labor force participation impact on mothers who, without the policy, would have stayed home even when the child entered school. Negative effects on parents well-being could therefore persist because of increased difficulties balancing work and raising children even once the child enters school. Second, previous research documented negative effects on children’s developmental scores, health, and behavioral outcomes (e.g. Baker, Gruber, and Milligan 2008, Kottelenberg and Lehrer 2014). If these effects persist once they enter school, then parents’ health and behaviors could also be negatively impacted since the well-being of the child and the well-being of parents are linked. However, recent work by Haeck, Lebihan, and Merrigan (2018) rule this option out. They show that the childcare reform had limited negative effects on children’s health and behaviors once children are in primary school or even high school. More precisely, they show that for children in school, only the impact on children’s emotional disorder and anxiety persists but the magnitude of the effect is considerably smaller than the effect measured for preschoolers. Since children are doing relatively better once in school, parents may also experience a similar improvement. Third, the negative effects on parents may simply persist through some kind of auto-regressive process. Finally, a fourth possible channel could be that parent-child interactions become less stressful when the child enters school. Parents may be less stressed and anxious when their child becomes older because they are more mature and independent, and require less effort and time investment from parents.

For our study, we use the National Longitudinal Survey of Children and Youth (NLSCY), cycles 1 (1994–95) to 8 (2008–09). This survey is a representative sample of the Canadian population of children and their families. To estimate the policy effects, we rely on a non-experimental evaluation framework with multiple pre- and post-treatment periods. Effectively, we compare Quebec parents before and after the reform to comparable parents in the rest of Canada (RofC). Our empirical strategy allows us to differentiate the intensity of treatment for each cohort, given the gradual implementation of the policy.

Our estimates suggest that, overall, the reform had negative effects on the mental health and behaviors of parents with preschool children. However, for some outcomes, our estimates by wave suggest that the effects decrease over time and eventually become statistically insignificant by 2008. We then extend the analysis to parents with at least one child aged 5–9 years old and no children aged less than 5. We show that, once children enter school, the negative effects of the program on parental mental health and behaviors fade away, except for the negative effect on mother-child interactions. Finally, Lefebvre, Merrigan, and Verstraete (2009) show that the reform had long lasting positive labor supply effects (once the child enters school), especially for less educated mothers, yet the effects we document do not differ by maternal education. Increased labor force participation for mothers. may explain the negative effect on positive interactions, a measure of positive quality time spent with the child.

The outline of the paper is as follows. Section 2 describes Quebec’s childcare reform. Section 3 describes the data set, while Section 4 presents the methodology. Empirical results are presented in Section 5, and Section 6 concludes the paper.

2 Quebec’s Childcare Reform

In the late 1990s, the government of Quebec initiated the gradual implementation of a low-fee childcare network for children under 6 years of age. The low-fee childcare spaces could be purchased at a single price of $5 per day per child. On September 1, 1997, only the 4-year-olds were eligible for low-fee spaces. They were followed by the 3-year-olds on September 1, 1998 and the 2-year-olds on September 1, 1999. By September 1, 2000, all children aged less than 59 months, not entitled to kindergarten because their fifth birthday occurs after September 30, became eligible for subsidized childcare. While all children were eligible, the number of available spaces at the time did not meet demand. Between 2000 and 2012, the number of low-fee spaces increased from 85,000 to 217,000 spaces, thereby releasing the capacity constraint. The price of low-fee childcare increased from $5 to $7 per day per child in 2004. Overall the total number of regulated childcare places in Quebec rose from 78,864 in 1997 to 258,366 in 2013, and the total government subsidy reached 2.3 billion dollars for fiscal year 2012–2013 (Conseil du Trésor-Quebec, Budget 2012–2013). In contrast, the number of subsidized childcare spaces in the other Canadian provinces was relatively small compared with the province of Quebec and changed little between 1997 and 2009 (Haeck, Lefebvre, and Merrigan 2015). This reform drastically changed the way in which preschool children were cared for in Quebec, while no comparable changes were observed in the rest of Canada.^[5]

Figure 1 presents the mean hours (conditional and not conditional on the use of childcare) per week that children aged 1 to 4 spent in their primary care arrangement, but also the labor force participation of mothers (two-parent and single-parent families) and fathers (in two-parent families) for these children in Quebec and the RofC. Haeck, Lefebvre, and Merrigan (2015) show that not only did a much larger percentage of infants and toddlers start to attend daycare in Quebec following the reform, but that the intensity of care also increased. Relative to the labor force participation, the main changes are for mothers. Indeed, for two-parent families in Quebec, mothers’ labor supply increased in most waves, starting at 55 percentage points in 1994 and reaching 76 percentage points in 2008. In contrast, no significant changes in the hours of care and maternal labor force participation has occurred in the RofC. For single mothers, there are large increases in labor force participation for both Quebec and the RofC, but the original gap, in favor of mothers in the RofC is totally closed by 2008.

Figure 1:

Trends in childcare hours and parental employment in Quebec and the rest of Canada, 1994–2008.

Shows the evolution of the mean number of hours per week spent in the primary mode of care in the Rest of Canada (left panel) and Québec (right panel) non conditionally (hollow square) and conditionally on attending childcare (hollow circle). The sample includes NLSCY cross-sectional children aged 1 to 4. Source: Haeck, Lefebvre and Merrigan (2015).

Along with subsidised childcare, the policy implemented changes for school-age children. First, full-day kindergarten replaced half-day kindergarten for 5-year-olds in school as of September 1998.^[6] Haeck, Lefebvre, and Merrigan (2015) show that the kindergarten policy itself did not have any impact on mothers’ labor force participation, but the combination of the low-fee daycare program and full-day kindergarten did. Second, before- and after-school daycare were now also offered to children aged 5 to 12 on school premises, also at the low-fee of $5 per day per child and $7 as of 2004. Lefebvre, Merrigan, and Verstraete (2009) and Haeck, Lebihan, and Merrigan (2018) show, respectively, that the before- and after-school program had no effect on labor supply and child outcomes. Indeed, these studies show that the price change was much smaller and could only marginally affect mothers’ labor supply decisions and child well-being. Clearly, this past research demonstrates that the main impacts on children and mothers are caused by the childcare policy.

3 Data

To estimate the long-term impacts of the reform on parents, we use the National Longitudinal Survey Children and Youth (NLSCY). The NLSCY contains both child and parental outcomes as well as extensive variables related to parental labor supply, childcare use, and other demographic characteristics. The survey started in 1994–1995 (wave 1) and ended in 2008–2009 (wave 8). This implies that we can observe families with young children in Canada four years prior to the implementation of the reform and for more than 10 years after. The first longitudinal cohort contained approximately 22,000 children aged 0 to 11 in 1994–1995. These children were originally scheduled to be surveyed every two years. In 1996, however, approximately 6000 children were dropped from the survey. The remaining children were followed until 2007–2008, the last cycle of the survey. As of cycle 2, in 1996–97, in addition to the original cohort, a new cohort of 0 to 1-year-olds was selected in every cycle. These cohorts were surveyed until the age of 5 and in some cases until the age of 7 or even 9. Therefore, we can observe families with children aged 0–9 years old throughout the length of the survey. In the NLSCY, all outcomes are reported by the person most knowledgeable about the child (almost always the mother).

We start by replicating earlier results on families with preschoolers, but over a longer observation period, as the program matures. In particular, we test whether the adverse effects of the program persist after 2006, when efforts around the qualification of educators and improvement in their salaries were undertaken by the government and the number of spaces in the network stabilized (Haeck, Lefebvre, and Merrigan 2015; Haeck, Lebihan, and Merrigan 2018). We then determine whether the effects on families with preschoolers persist when the policy is no longer contemporaneously effective, that is, when children are in school (ages 5–9). For children in school, we only retain in our sample families with no children of preschool age in order to avoid any confounding effects of the program.

Given the gradual implementation of the reform, children and parents were treated differently by the policy over the years. Table 1 shows the various treatment groups by presenting the eligibility of children according to their age and NLSCY wave from which they are sampled. The grey-shaded area highlights the post-reform years while the unshaded area refers to the pre-reform years. Numbers indicate the number of years of eligibility for subsidized childcare. To calculate the number of eligible years, we always use December 31 of the first year of the two-year period as a reference.^[7] For example, for wave 4 (2000–01), the reference point is the child’s age on December 31, 2000. For the index 0.5, the child is eligible for a few months, not a year. Table 1 also shows that some age groups are not observed in certain waves. For example, parents with children aged 8 and 9 years old are not observed in 2004 (wave 6). So, given the structure of NLSCY, school-age children are split in two groups: the 5- to 7-year-olds and the 8- to 9-year-olds.^[8] We exclude wave 3 data for children 0–7 years old in order to avoid the overlap of treated and untreated children in the same wave.^[9] Moreover, the number of regulated childcare spaces did not change in the early years of the reform (before 1999). At that time, existing spaces were converted to low-fee spaces (see Figure A.1 in Haeck, Lefebvre, and Merrigan 2015). Nevertheless, we assess the robustness of our results to this restriction in the empirical section.

Clearly, the number of years in low-fee childcare increased over time (Table 1). As such, families with children aged 0–4 years in Baker, Gruber, and Milligan (2008) were only treated for a few months to 2 years (waves 4 and 5 of the NLSCY).^[10] Here, we add an additional six years of data and we also observe school-age children. This allows us to study the impact of the reform on parents with preschool children over a longer period. It also enables us to analyze the long-term effects of the policy on families with school-age children (aged 5–9 years old). In both groups, we can now observe families with children who were eligible for low-fee childcare since birth and were therefore highly exposed to the reform (up to 5 years of treatment).

Table A.1 summarizes the various treatments by age group while presenting the eligibility of families according to child age and NLSCY wave. We distinguish four cohorts defined by their eligibility status to low-fee childcare: (1) not eligible; (2) partially eligible, with the estimated effects for this group symbolized by β ^P ; (3) fully eligible but observed in the early years of the program, β ^FE ; and (4) fully eligible and observed in the later years of the program, β ^FL . Families never eligible for the childcare reform are in the first cohort. In the NLSCY, this cohort includes families with children aged 0–7 years old in waves 1 and 2 and families with children aged 8–9 years old in waves 1 to 4. The second cohort includes families partially treated by the reform; these families were all eligible for low-fee childcare but only for a maximum of 3 years. As mentioned earlier, waves including families with children who were not treated and children who were, are excluded from our main sample. The third cohort includes families with children who were treated since birth, but were still in the early years of the program. Finally, the fourth cohort includes families with children who were eligible since birth and were observed in the later years of the program. Table A.1 reveals that, for the 8- to 9-year-olds, we only observe families with children who were partially treated. For the 5- to 7-year-olds, we observe families with children who were fully treated but only in the early years of the program. Finally, as to fully treated children in the later waves of the program, we only observe families with children aged 0–5 years old. As suggested by Haeck, Lebihan, and Merrigan (2018), we consider the fact that families are exposed very differently to the program. Not separating families with different treatment intensity results in estimates that potentially mask important differences. In this study, we compare families with similar exposures to the program (partial vs. full exposure). As will become clear in the next sections, the timing and duration of eligibility should always be kept in mind when interpreting the results on the effects of the reform.

Thus, given the data availability and eligibility for subsidized childcare depending on the age of the child, we focus our analysis on the parents with children aged 1–9 years old. The evaluation is performed for three separate age groups: parents with 1- to 5-year-olds not in school; parents with 5- to 7-year-olds in school (and no children of preschool age); and parents with 8- to 9-year-olds (and no children of preschool age). We exclude from our samples children 12 months old or less who generally do not attend childcare since parental leave is 50 weeks long in Canada.^[11] We also include parents with 5-year-olds not in school in the sample of preschool children as they are likely to be in subsidized childcare before being eligible for kindergarten.^[12]

Following Baker, Gruber, and Milligan (2008), Kottelenberg and Lehrer (2013, 2014, and Haeck, Lebihan, and Merrigan (2018), we focus on two-parent families to avoid interference with other policies targeting low-income families (largely represented by single-parent families). Various provincial and federal reforms have been implemented since 1997 and could interact with the low-fee childcare reform. Baker, Gruber, and Milligan (2008) and Milligan and Stabile (2007) show that changes in family/child benefits have a statistically significant and relatively large impact on different outcomes for single-parent families, but little impact on two-parent families. In addition, the Government of Quebec introduced a new work incentive policy in 2005 to support and develop the work effort of low-wage workers as well as encourage people to exit welfare into work (Ministry of Finance of Quebec 2004). Therefore, since any specific policy shock in Quebec coinciding with the universal childcare reform may bias our results, we focus on two-parent families.^[13]

We measure the impact of the reform on parental health and parental behavior. For parents’ health, we use the following outcomes: (1) the mother’s health status (1: excellent, 0: not excellent); (2) the father’s health status (1: excellent, 0: not excellent); and (3) the mother’s depression score (score ranging from 0 to 36). A high score indicates the presence of symptoms of depression. All questions on parents’ health are asked to households with children aged 0 to 9. For parental behavior and parenting per se, several measures are available: (1) the family dysfunction index (score ranging from 0 to 36); (2) the positive interaction score (ranging from 0 to 20); (3) the hostile/ineffective parenting score (ranging from 0 to 25); (4) the consistent parenting score (ranging from 0 to 20); and (5) the aversive parenting score (ranging from 0 to 20). A high score for (2) and (4) indicates positive parental behavior for child well-being while the opposite is true for (1), (3), and (5). The questions on parents’ behavior are asked when children are 2–9 years old, except for the family dysfunction score, which is available for all parents. The subquestions used for each measure are reported in Table A.2.^[14] We construct our outcomes in the same way as Baker, Gruber, and Milligan (2008) and Kottelenberg and Lehrer (2013, 2014 in order to compare our results with these studies.

We also use the same control variables as Baker, Gruber, and Milligan (2008), Kottelenberg and Leher (2013), and Haeck, Lebihan, and Merrigan (2018) in our regression analysis to ensure that any differences between our results and theirs is not due to controls or methods. The control variables are: the sex of the child, the age group of the mother and father at the child’s birth (14–24 years old (omitted), 25–29, 30–34, 35 or more), the mother’s and father’s highest level of education (less than a high school diploma, high school diploma, some post-secondary education, with post-secondary diploma (omitted)), a dummy for whether or not the mother or father was born in Canada, the size of the area of residence (five groups from rural population to 500,000 residents or more (omitted)), the presence of older children (no older child (omitted), one older child, at least two older children), the presence of younger children (no younger child (omitted), one younger child, at least two younger children), the presence of children of the same age, and dummies for the age of the child. Summary statistics for parents with children aged 1–9 years in Quebec and the RofC pre- and post-reform are presented in Table A.3.

4 Empirical Strategy

To estimate the long-term effects of the daycare reform, we rely on a non-experimental evaluation framework based on multiple pre- and post-treatment periods (difference-in-differences model). The treatment group includes Quebec parents with children of a given age before and after the reform. The control group includes parents in other Canadian provinces with children of the same age observed during the same time period. Only Quebec parents are affected by the reform and periods of pre- and post-treatment depend on the age of the child (see Table 1). To account for the gradual implementation of the reform, we allow the effects of treatment to differ in each of the post-reform waves. The empirical model is as follows:

where Y _ipt represents a parent outcome for child i in wave t in province p. Regional and temporal differences are captured using province fixed-effects (θ _p ) and survey wave^[15] fixed-effects (ω _t ). To capture the gradual impact of the reform due to the phase-in by age group, we estimate the coefficients on the interaction between wave dummies (W _t ) and a Quebec province dummy (Q _it ). The term Q _it is a dummy variable taking the value 1 if child i lives in Quebec in wave t and 0 otherwise. The term W _t takes the value of 1 if the wave is greater than or equal to T = 4 for families with children 1–7 years old and T = 5 for families with children 8–9 years old (see Table 1). The term X _it is a vector of socioeconomic control variables and ɛ _ipt is an error term.

Standard errors are estimated using the 1000 bootstrap weights provided by Statistics Canada. This procedure accounts for the complex survey design of the NLSCY. Since we estimate impacts for multiple outcomes simultaneously, we also adjust our p-values following Simes (1986). This correction assumes that our outcomes are correlated with one another and avoids the possibility of over-rejecting the null hypothesis when studying multiple correlated outcomes. As suggested by Shaffer (1995), our adjusted p-values are computed by sub-groups (health; behavior).^[16]

Our empirical strategy relies on two critical assumptions. First, our approach assumes no selection based on province-specific transitory shocks. Second, in the absence of the policy, mean outcomes of Quebec families with children would have followed a similar trend as those of comparable families in the other provinces.

Concerning the first assumption, when the reform was announced, ineligible children were already born. Parents could not have delayed conception to be eligible to subsidized daycare. Also, although parents outside of Quebec could have moved to Quebec to benefit from the childcare reform, migration data do not support this idea (see Lefebvre, Merrigan, and Verstraete (2009) for more details). Another reason that could influence group composition relates to the ferility decision of parents. However, relative fertility rates did not change immediately after the reform. Figure A.1 shows that the fertility rate in Quebec since 1990 has been very similar to the Canadian fertility rate. We observe a slight change as of 2007, at the very end of our observation period, not enough to impact our results on children age 1 and over.

Concerning the second assumption, we cannot observe untreated families in Quebec post-reform, but we can observe trends in the outcome variables in the treatment and control groups prior to the reform. Figure 2 shows the evolution of a few outcome variables pre- and post-treatment. The paths of these outcomes in Quebec and the RofC are similar, especially when we have multiple years of observation pre-reform (the 8–9 years old), prior to the introduction of the policy, regardless of the age of the child. This only provides visual information for two of our eight outcomes. Following the presentation of our main results below, we formally test the common trend assumption for all outcomes and interpret our results in light of these findings.

Figure 2:

Mean values of measures for parental outcomes by region and child age: waves 1–8 of the NLSCY.

Shows the trajectories for the mean of two outcomes (standardized) by age of the child for Quebec and the Rest of Canada (RofC). For families with children in school, figures are obtained for families with the youngest child aged 5-7 years (in school) or 8-9 years.

Data: NLSCY.

We also formally test the common trend assumption for all of our outcomes and age groups using an event-study analysis (Tables A.4 and A.5). The reference category is always wave 1. Results suggest that for preschoolers, the common trend assumption holds for all outcomes: the null hypothesis cannot be rejected for β ₂ and β ₃. For children aged 5 to 7 the common trend assumption generally holds as well, except for positive interaction and hostile parenting in wave 3 (β ₃). It also does not hold for positive interaction for children aged 8 to 9 in wave 4 (β ₄). This suggests that our control groups are generally reliable, except when studying the impact of the reform on positive interaction.

5 Econometric Results

We start by providing evidence on the effects of the program on the well-being of families with preschool children. These include the families with 1- to 4-year-olds and also the 5-year-olds not yet in school. Since the policy was implemented progressively, relative to former studies, we take into account the differences in intensity of treatment. Some families were treated partially while others fully. We also assess whether the initial impact on families with preschool children persisted as the program matured, especially as of 2006.

We then present the results once these children enter school. We estimate the effects of the program on the families with children aged 5–7 years old and 8–9 years old (without children of preschool ages).

This allows us to estimate the contemporary effect of the reform on parents with preschool children, and the spillover effects into the school years while taking into account the intensity of treatment. We present the average effect over the entire post-reform period in the tables, along with the estimates per wave. We also report whether families with children were treated partially (β ^P), full early (β ^FE), or full late (β ^FL), based on Tables 1 and A.1. We report the coefficients, standard errors, and results from several statistical tests. Estimated coefficients that are statistically significant based on adjusted p-values are presented in bold.^[17] We also report a plus or minus sign for each outcome showing the direction the effect must take for the policy to be beneficial to families. To ease the interpretation of our results, all non-binary outcomes were standardized for all respondents to have a mean of 0 and a standard deviation (SD) of 1. The coefficients can thus be interpreted in terms of changes in SDs. We focus on the estimates in boldface.^[18]

Table 2 presents the estimated effects of the subsidized childcare policy on parents’ health and behavior when their child is 1–5 years old. The results for parents with children in school, aged 5 to 7 and 8–9 years old, are presented in Table 3.

6 Conclusion

Our paper shows that the Quebec childcare policy had detrimental effects on parental mental health and behavior for parents of preschoolers. However, the effects documented early on in the reform on parents of children not yet in school appear to diminish over survey cycles for positive interaction and consistent parenting, but less so for positive interaction. Effects remain persistent for hostile parenting and aversive parenting. Also, our results suggest that once children enter school and the parents no longer have preschool children in the household, most detrimental effects fade in later survey cycles.

We are unable to identify the exact drivers of these results, but evidence presented in previous studies of this reform suggests a few possibilities. First, as mentioned earlier, previous research documented that the childcare reform increased maternal labor force participation, work hours and children’s time spent in childcare. The effects on maternal work were documented both in the preschool years and once children were in school. These effects mechanically lower the number of hours available for parents to interact with their child. Second, childcare quality was considered fairly low, which may have had an impact on parental behavior. Third, negative effects on preschoolers also appear to fade once children enter school. Parental behavior and mental health therefore seem to mirror what has been previously documented in children.

One of the limits of our study is that while the frequency of parental interactions may have decreased following the reform, the overall quality of the interactions for a given quantity of time spent with the child is not directly captured by our measures. Another limit relates to the lack of data after 2008 which limits our ability to follow children in more recent years of the program.

Our results, in conjunction with those of previous studies, suggest that the negative effects of the policy on preschool children and their parents should be of concern for policy makers seeking to provide universal care to children. However, concerns about the long run effects on parents and children once children enter school should be limited given the evidence presented here and in Haeck, Lebihan, and Merrigan (2018). Any major policy that seeks to radically increase the labor force participation of mothers with young children through highly subsidized childcare must thoroughly consider all family dimensions, in particular physical and behavioral, before its implementation. Finally, changes in childcare subsidies may affect child and parental well-being through increased household income, enlarging consumption possibilities. The impact of the Quebec childcare reform on household finances and consumption need to be further studied in the future.