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Intra-household allocation of family resources and birth order: evidence from France using siblings data

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Abstract

We examine the effect of birth order on education, occupation, and parental transfers using four cross sections of the French Wealth surveys conducted between 1992 and 2010. Estimates from ordered models confirm the presence of a first born advantage in education and occupation, the latter persisting to a lesser extent after controlling for education. Strikingly, parents are on average more likely to make transfers to first-born children, although the vast majority provides cash or property gifts to all of their children. This first-born advantage in transfers is uncorrelated with the likelihood of having attained a higher education or better occupation. Overall, our findings suggest that in France, the mechanism supporting the first born advantage may not stem from confluence effects or family resource dilution.

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Notes

  1. There are also reasons why the last born could, in theory, be favored, e.g., parental earnings increase over the life cycle (hence, there would be fewer resources for the first-born children if households face liquidity constraints) and older mothers are more experienced.

  2. Databases are available free of charge for use in research from the French Data Archives for social sciences (Réseau Quetelet, http://www.reseau-quetelet.cnrs.fr/spip/?lang=en).

  3. The number of respondents is, respectively, 9530 in 1992, 10,168 in 1998, 9692 in 2004, and 15,006 in 2010. In the 2010 Wealth survey, the sample includes 2218 households living in overseas departments. These households were excluded for comparability purposes over the period.

  4. Although we can always identify recomposed families, we treat co-resident half siblings as pure siblings since we do not know which siblings are not those biological children of the reference person. All our results hold when removing recomposed families from the sample.

  5. In the Wealth surveys, there is no information on education for children under 17.

  6. Also, we choose to exclude the few observations (N = 413) in which the household head was under 45.

  7. In 2004, the questionnaire did not include any information on education and occupation of children living with their parents. This explains the smaller number of observations for the 2004 survey.

  8. However, there is no information on the amounts received by children.

  9. Among workers, the male-female ratio was 2.5 for cohorts born before 1945, 4 for cohorts born before 1955 and 1959, 6 for cohorts born between 1975 and 1979, and even 8 for cohorts born since 1980.

  10. Regional dummies and dummies for size of urban unit, which are expected to explain differences in educational supply, will also be taken into account.

  11. However, compared to two-child families, the proportion of having high education is lower for one-child families (42.3 % compared to 45.6 %).

  12. For families with four or five children, the proportion of children with more than high school education appears to be U-shaped with respect to birth order in the USA (Hanushek 1992). Kantarevic and Mechoulan (2006) analyze this puzzling stylized fact and find that it stems from the confounding factor of the mother’s age at birth. That is, when age of the mother at birth is controlled for, the inverse U-shape relationship disappears.

  13. This includes for instance the degree of parental altruism which should be positively correlated with investment in the human capital of children (Becker and Tomes 1986).

  14. We set μ 0 = −∞ and μ K  = +∞ and assume that the thresholds are strictly ordered (μ k−1 < μ k ).

  15. For co-resident children, the 1992 Wealth survey indicates whether they have completed the undergraduate level or the graduate/postgraduate level.

  16. We have also estimated ordered regressions using the birth order index proposed by Booth and Kee (2009). These additional results, which are available upon request, lead to similar conclusions.

  17. In our regressions, we control for the head’s age at birth since we do not always have information on both spouses for each child. Recall that the parent sample is constructed by assembling information on the head’s characteristics and spousal characteristics, if any.

  18. We have also considered a piecewise linear function for the child’s age, by adding both age and birth cohort dummies interacted by age in the regression.

  19. We have also estimated the random effect ordered Probit regression on the subsample of families whose youngest child is at least 24. The corresponding estimates are very close to those reported in column 1 of Table 3, with a negative and significant correlation between birth order and education.

  20. By definition, parental characteristics which remain constant at the sibship level are picked up by the family fixed effect and are thus excluded from the regression.

  21. We thank an anonymous referee for this insight.

  22. We also estimated our regressions on the subsample of families where all siblings are employed and reach similar conclusions concerning the role of birth order. We also investigated the relationship between unemployment and family characteristics, with a focus on non-co-resident children interviewed either in 1998, 2004, or 2010 given data constraints. When estimate a random effect Probit model to explain the probability for a child to be unemployed, we find no significant relationship between the probability of being unemployed and birth order.

  23. For three children and higher families, results from Wald tests show that the birth order coefficients are not jointly significant.

  24. At first sight, it could be a little surprising that the random and fixed effect estimates differ since there should be no correlation between birth order and family-level unobservables conditional on family size. However, both regressions are not estimated on the same sample of children. The fixed effect ordered models rely on the estimate of conditional Logit models, so that sibships in which children all have the same occupation are excluded.

  25. Specifically, these authors find that first borns get more education but also argue that by age 30, no birth rank effect persists across siblings. They explain their findings by claiming that later borns are more risk taking than first borns. It remains to be explained why later borns would not outpace first borns after age 30, however.

  26. Hence, the difference in transfer income derivatives has to be equal to minus one, a property called redistributive neutrality (see Laferrère and Wolff 2006).

  27. Alternative explanations have been suggested to explain equal sharing within the family. For instance, parents may suffer from a psychic cost when deviating from an equal allocation of resources (Wilhelm 1996).

  28. In the French Wealth surveys, there is no information on the amount per transfer for regular or irregular cash gifts.

  29. Assuming that the probability of a child to be in financial problems is p, then the probability of transferring to all children is p n for a family with n children. The probability of an uneven distribution is 1 − p n − (1 − p)n, which is an increasing function of n.

  30. Children aged between 30 and 39 are more likely to receive a gift from their parents, while the other characteristics have no significant influence (presumably due to small sample size).

  31. These additional results are available upon request. In the fixed effect linear regression, both gender and education of the child have no significant influence on receiving money from parents.

  32. A few papers have investigated the existence of interactions between siblings in long-term care decisions (Heidemann and Stern 1999; Engers and Stern 2002; Byrne et al. 2009; Fontaine et al. 2009).

  33. Data limitations do not allow us to go further. In particular, we would like to know the amounts at stake to quantify the transfer differentials and compare those to other siblings’ outcome differentials. While we can only observe the presence of financial transfers, we are not able to compare amounts as well as other forms of cash gifts across siblings.

  34. At the same time, there may be strategic considerations within the sibship to avoid the burden of providing care to parents. See in particular Konrad et al. (2002) and Stern (2014).

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Acknowledgment

We are indebted to two anonymous reviewers for their very helpful comments and suggestions on previous versions of this paper. We also thank Hippolyte d’Albis, Carole Bonnet, Laurent Gobillon, Robert Pollak, Anne Solaz, and seminar participants at the Paris Seminar in Economic Demography, McMaster University, the University of Guelph, Ryerson University, and L’Université du Québec in Montréal. Any remaining errors are ours.

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Authors and Affiliations

  1. School of Public Administration, Dalhousie University, Nova Scotia, Canada

    Stéphane Mechoulan

  2. LEMNA, Université de Nantes, BP 52231 chemin de la Censive du Tertre, 44322, Nantes Cedex 3, France

    François-Charles Wolff

  3. INED, Paris, France

    François-Charles Wolff

Authors
  1. Stéphane Mechoulan
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  2. François-Charles Wolff
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Correspondence to François-Charles Wolff.

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Responsible Editor: Responsible Editor: Junsen Zhang

Appendix

Appendix

Table 7 Ordered Probit estimates for education and occupation, by family size
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Mechoulan, S., Wolff, FC. Intra-household allocation of family resources and birth order: evidence from France using siblings data. J Popul Econ 28, 937–964 (2015). https://doi.org/10.1007/s00148-015-0556-x

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