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The effects of the Chilean divorce law on women’s first birth decisions

Abstract

In 2004 a new law in Chile allowed couples to divorce for the first time. The law also established compensation in case of divorce where one spouse sacrifices professional development or advancement for the good of the household. Using birth histories constructed from the Chilean Social Security Survey (Encuesta de Prevision Social–EPS) Panel 2002–2009, we investigate the effect of the divorce law on a woman’s decision of when to have a first child. We find that the divorce law increases the hazard rate of having the first child by 61 % for more educated women, controlling for socioeconomic characteristics, length of marriage and the negative trend in fertility rates observed in Chile since the mid-1960s. We also find that a given percentage increase in a woman’s potential income will increase the hazard rate by a greater percentage increase after the passage of the law.

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Fig. 1
Fig. 2

Notes

  1. The amount of compensation is determined by a court, which takes into account the beneficiary’s age, level of education, length of marriage, and number of children, as well as other socio-demographic characteristics.

  2. Chilean Library of Congress. http://www.bcn.cl/carpeta_temas/temas_portada.2005-10-27.7388460505.

  3. Separation provided a woman with a food pension, but there is evidence that these pensions are frequently not paid: administrative data shows that 80 % of petitions to the courts concerning children’s issues are related to nonpayment of a food pension. See Senate of Chile, Records of Session 12 (15 July 2003).

  4. We would like to thank an anonymous referee for this argument.

  5. These figures are provided by the National Institute of Statistics of Chile (INE). The INE produces the Vital Statistics Report (Reporte de Estadísticas Vitales) annually, using administrative information on births, marriages and other variables collected from birth registry offices. In Fig. 1 the number of marriages for the years 2001, 2002 and 2004 is estimated using the annualized growth rate of first marriages for the periods 2000–2003 and 2003–2005, respectively. The number of first births for the years 2001 and 2002 is also estimated using the annualized growth rate of the number of first births for the period 2000–2003.

  6. The administrative data do not distinguish the average ages at first marriage of women who have children and women who do not.

  7. In general, between 1998 and 2009, the 15 countries of the Eurozone decreased by 0.8 %. http://ec.europa.eu/eurostat.

  8. In 2013, a benefit scheme was introduced to provide financial support to couples with more than two children.

  9. In 2005, new domestic violence legislation was introduced. However, there is no clear evidence of the effect of domestic violence on fertility decisions. Moreover, to our understanding, there are no exclusive social programs targeting single mothers.

  10. Peters (1986, 1992), Allen (1992), Zelder (1993), Weiss and Willis (1997), Friedberg (1998) and Wolfers (2006) analyze the effects of divorce laws on divorce rates. Regarding the relationship between divorce laws and women's labor supply, the literature focuses on the response of women's labor supply after divorce: Johnson and Skinner (1986), Gray (1998) and Stevenson (2007). Regarding human capital accumulation, King (1982) and Stevenson (2007) analyze the effect of divorce laws on couples’ investments in the education of spouses and children, as well as household specialization.

  11. Gruber (2004) provides a summary of when these laws were implemented in each state.

  12. CIA World Factbook. https://www.cia.gov/library/publications/the-world-factbook/geos/ci.html.

  13. To our knowledge, the only study that examines the Chilean DL is Heggeness (2009), which analyzes the law’s effect on educational enrollment using a difference-in-difference approach.

  14. To see this, we compute the difference between the share of first births for more educated women (those with more than 12 years of education) and the share for women with 10–12 years of education for each year after the DL was passed in 2004. We compare these differences with the corresponding difference for the year 2003, which is one year before the DL was passed. The formula is \(DD_{13}^{y} = \left( {f_{13}^{y} - f_{10\,to\,12}^{y} } \right) - \left( {f_{13}^{03} - f_{10\,to\,12}^{03} } \right)\) for y from 2005 to 2008. The difference-in-differences increases from 0.77 % in 2005 to 3.06 % in 2008 (see Fig. 2). The result is unchanged if women 19 or younger are excluded from the analysis.

  15. The original sample was collected in 2002. The 2004 wave added around 3000 new individuals to the sample, while the 2006 and 2009 waves update the individuals' information.

  16. The results are conditional on not having conceived before the age of 18. In the administrative data, the percentage of women conceiving their first child at age 18 or before is very small. The calculations presented in the text are based on administrative data. According to the Ministry of Health of Chile (2013) the percentage of conceptions for women under the age of 19 has remained relatively flat from 2004 to 2011 at 15.5 % (there are no figures for first conceptions alone). In the administrative data there were substantially fewer first conceptions for this age group, and their exclusion did not significantly affect the difference-in-difference analysis, and the average age at first birth in Fig. 2.

  17. Less than 3 % of women in the sample are married or cohabitating before reaching 18 years of age. In such cases, we assume that their relationship started at age 18. Before 2009 the survey does not collect information on the timing of the current marriage or cohabitation. Therefore the exact duration of a marriage or cohabitating relationship may be mismeasured when more than one occurs. Less than 18 % of the women in the sample report having terminated a first marriage or cohabitating relationship, and only 11.5 % of non-single women report more than one marital or cohabitating relationship.

  18. The variables in W are the level of education level of the woman’s father and mother, an indicator variable for whether the woman married before 2004, and a set of indicator variables to control for the region of residence of the woman when first interviewed.

  19. The actual final level of education is not available in the EPS.

  20. Specifically we assume the following: (a) women who last report primary or secondary education will complete that education level; and (b) women who last report enrollment in the final year of secondary education or who report attending post-secondary education in every wave of the EPS will complete the corresponding track of post-secondary studies.

  21. We would like to thank an anonymous referee for suggesting these robustness checks.

  22. To test for anticipation effects, we substitute the time-varying covariate that goes from 0 to 1 to indicate the implementation of the DL in November 2004, for a duration spline. This duration spline enters the hazard in 1999, when the DL was presented to the congress. We also include nodes in May and November of 2004 for the approval and implementation of the DL. The results (not shown in the paper) suggest that the DL has no anticipation effect on the hazard of having a first child. Indeed, this result is not surprising, given that the DL is the last of multiple attempts to introduce divorce in Chile. Moreover, none of the results in this study change substantively when the indicators for region of residence are replaced with the woman’s region of birth.

  23. Ermisch and Pronzato (2010) show that an additional year of parent’s education increases their children’s education by at least one tenth of a year.

  24. We have also included a piece-wise linear time trends with knots at every 10 years after the start of the sample. The knot at 1990 is critical since it represents the start of the post Pinochet period. The knot at 2000 coincides with a presidential election. Moving the knot from the year 2000 to non-election years increase the variability of the estimate associated with the DL, and hurts our significance levels in the regression where the highest education level is some university experience, but not in the case where the highest education level is university degree.

  25. See Heckman (1979).

  26. The standard errors for the estimates of the interaction of the DL variable with the woman’s potential income have not been adjusted to take into account the fact that the woman’s potential income is a generated regressor.

  27. Column (3) of Table 6 shows the full set of results for Eq. (3).

  28. As an alternative specification, we include in Eq. (3) the interaction of DL with the woman's expected education (E·DL(t)). The estimates from this specification corroborate the negative effect of the woman’s potential income on the hazard of having a first child.

  29. Our regressions do not control for general economic conditions. A sustained economic growth in Chile starting in 1982 has resulted in low levels of unemployment and improvement in women’s educational attainment, which we do control for. Therefore it is unlikely that fertility decisions are negatively affected by deteriorating economic conditions.

  30. We use a Mincerian equation to estimate the partner’s expected income. Using the EPS 2004 male sample, we regress the natural logarithm of hourly income in 2004 on education, labor force experience, and its square. The equation is used to predict log hourly income at age 25. The EPS provides information about the partner’s education only in the marriage histories updated after the EPS 2006 (in earlier waves this information is not collected). However, for a marriage or cohabiting relationship starting before 2004, it is still possible to recover the partner’s education from the household members’ demographic characteristics if a married woman reports only one marriage (58 % of the sample). We mean-fill by woman’s education in the case of missing values.

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Acknowledgments

We would like to thank Susan Gensemer, Alfonso Flores-Lagunes, Pieter Serneels, two anonymous referees for their useful comments and the Subsecretary of Social Security of Chile for providing the data. We are also thankful to the participants at the Gender Economics session at the 2012 Meeting of the Latin American and Caribbean Economic Association. All errors are our own.

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Correspondence to Jose V. Gallegos.

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Gallegos, J.V., Ondrich, J.I. The effects of the Chilean divorce law on women’s first birth decisions. Rev Econ Household 15, 857–877 (2017). https://doi.org/10.1007/s11150-015-9307-8

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Keywords

  • Divorce law
  • Time to first birth
  • Chile

JEL Classification

  • J12
  • J13