Abstract
An intergenerational model is developed, nesting heritable earning abilities and credit constraints limiting human capital investments in children. Estimates on a large, Finnish data panel indicate very low transmission from parental earnings, suggesting that the parameter of inherited earning ability is tiny. Family income, particularly during the phase of educating children, is shown to be much more important in shaping children’s lifetime earnings. This influence of parental incomes on children’s earnings rises as the children age because the returns to education rise. Despite Finland’s well-developed welfare state, persistence in economic status across generations is much higher than previously thought.
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Notes
Becker and Tomes (1979, 1986), Behrman et al. (1995). Easterlin et al. (1980) expressed the third alternative in terms of fertility outcomes, though it may be applied equally to quality rather than just quantity of children. Excellent surveys are provided in Behrman (1997), Solon (1999), Björklund and Jäntti (2009), Björklund and Salvanes (2010), and Black and Devereux (2010).
These correspond to the high-resource and low-resource family cases in Behrman et al. (1995).
In Eq. 5, the partial derivative of H c with respect to G c has the same sign as η, thus being negative (positive) for elasticities of substitution <1 (or >1).
Exploratory results in the present context suggest that the use of instrumental variables increases the estimated transmission from family income but not from parental earnings, though identifying appropriate instruments is, as always, a difficult task and this route is not followed in the present paper.
“A family consists of a married or cohabiting couple and their children living together; or a parent and his or her children living together; or a married or cohabiting couple without children. Persons living in the household-dwelling unit who are not members of the nuclear family are not included in the family population, even if they are related” (Statistics Finland 1995, p. 16).
Whether the adults with whom the child was living in a 1970 nuclear family are the biological parents is not known. Children recorded in the family unit comprise biological and adopted children of either spouse, though foster children and children in the care of the family are not classified as part of the family.
Among the children ages 0–16 in our database, less than three quarters of 1 % are not living with their family and hence omitted from the family-based sample. This very low fraction is in accord with a report by The Population Research Institute (1995), which shows that in the mid-1980s Finnish children continued to live with their parents even beyond age 16 more frequently than did children in the other Nordic countries: 95 % of boys ages 16–19 were still at home in Finland as were 88 % of girls.
“Income subject to state taxation does not include scholarships and grants received from the public corporations for study or research, earned income from abroad if the person has worked abroad for at least six months, part of the social security benefits received from the public sector and tax-exempt interest income” (Statistics Finland 1995, p. 18). Since 1985 taxable income in Finland does, however, include child, maternity, and unemployment benefits.
Most estimates of the intergenerational correlation in incomes are based on sample survey data, though a few studies have also extracted income measures from tax registers (Corak and Heisz 1999, on Canada; Österberg 2000, on Sweden; and Mazumder 2005, who uses the US Social Security Administration’s Summary Earnings Records). Register-based data, whether derived from employer or tax records, have some advantages over survey responses, particularly where the survey respondent is not the person employed. Yet register-based data also have some limitations. The Finnish data lack information on hours worked, for example, and hence are inadequate to examine hourly rates of pay and preclude computation of full income. The information on self-employment earnings also has some drawbacks. The usual caveats apply with respect to the role of capital income components within self-employment earnings. Moreover, given progressive taxes on each individual, an incentive exists to spread family self-employment income across family members where possible. Nonetheless, self-employment is not a major source of earnings for young people. For only 5.2 % of sons and 5 % of daughters are more than 90 % of their earnings, in an average year, derived from self-employment. For the remaining sons and daughters, self-employment earnings represent only 7 and 4.4 % of their earnings, respectively.
This literature also presumes a single child. An extension to multiple children is to be the subject of a separate paper, though preliminary findings indicate that the results in the present paper prove robust to including controls for family size.
Certainly, the majority of empirical studies, both of the USA and elsewhere, seek to relate sons’ earnings to earnings of their fathers. A much smaller number of studies correlate the earnings of daughters and of fathers, and a handful of studies have considered the mothers’ earnings. See Solon (1999), Tables 3, 4, 5 and 6 for a summary of estimates for both sons and daughters. Subsequent estimates of intergenerational income transmission for daughters include Österbacka (2001) and Chadwick and Solon (2002).
The appendices to this paper are available as Boston University Institute for Economic Development Discussion Paper No. 221 at http://www.bu.edu/econ/files/2011/09/Lucas_Kerr.pdf.
Source: Österbacka (2001Table 3 ). t statistics for a zero null hypothesis are shown in parentheses, calculated from the reported standard errors. In addition, Österbacka explores estimates within quintiles of parents’ earnings and the correlations between siblings’ earnings, while also undertaking a decomposition of the intergenerational correlations. See also Jäntti and Österbacka (How much of the variance in income can be attributed to family background? Empirical Evidence from Finland, unpublished).
See Tables 2 and 9 in Jäntti et al. (2007). t statistics for a zero null hypothesis are shown in parentheses, calculated from the reported confidence intervals.
The age range varies considerably over which earnings of children are included in prior studies. Solon (1992) imposes a lower bound at age 25 in his study of sons’ earnings, while the lower bound in Zimmerman (1992) is 29, and Dearden et al. (1997) look at UK sons and daughters when they are 33. Although only children who are at least 25 in 1985 are included in our initial analysis in Section 4, any earlier positive earnings (or wages) of these children are included in computing mean earnings, provided the son or daughter was at least 20 at the time of observation in the 1975 or 1980 census. To the role of age in these estimates, Section 5 returns.
Any distinction between this mean age measure, the mean age over which the individual is observed, or age at a specific point in time, is not always drawn in this context. However, in practice, the distinction has relatively little impact on the estimates.
Projecting to age 40 and censoring earnings data, Jäntti et al. (2007) obtain slightly higher estimates for sons. The following section returns to the issue of an age interaction.
Given the strong similarity in the results for earnings and wages in Table 2, for brevity, only results for the case of earnings are reported in the remainder of the paper. The specifications in family income include mean age of the father during periods when he reports positive income and a similar measure for the mother, both in polynomials through the fourth power, plus dummy variables for cases where no father or mother is initially present.
Several US studies also estimate transmission from family income to be larger than from parental earnings measures (see the review in Solon 1999). However, these studies uniformly concentrate upon transmission from parents’ family income to child’s family income, the latter presumably reflecting assortative mating in addition to the considerations in Section 1. As far as we are aware, no study incorporates both family income and parental earnings as regressors. A test for equality of all coefficients between the two stages of sampling within our data, based on the last specification in Table 3, gives F(17, 14,105) = 0.937 for sons and F(17, 13,611) = 0.564 for daughters. Pooling the subsamples appears not to be problematic.
Restricting the specification to be linear, as opposed to the far more parsimonious form, generates an F statistics for sons of F(97, 14,251) = 1.646 and for daughters F(97, 13,747) = 1.114. A least absolute deviation estimate at the median also generates only a slightly lower value for transmission from family income than does the estimate at the mean. See Electronic supplementary material—Appendix B.
Behrman and Taubman (1985) test a three-generational, schooling equivalent to Eq. 22 on a US sample. The same data set is used in Behrman and Taubman (1989) for a three-generational test of the Easterlin et al. (1980) fertility hypothesis. See also Robertson and Roy (1982), Warren and Hauser (1997), and Jeon and Shields (2005).
See the discussion in Card (1999) for instance.
Such an effect has previously been detected in the US context by Reville RT (Intertemporal and life cycle variation in measured intergenerational earnings mobility, unpublished), though Lee and Solon (2009) find less clear-cut patterns with respect to an age interaction. Equation 26 may be thought of as a specific form of the model developed in Haider and Solon (2006) which notes the potential age dependence of parameters such as β 1 and β 2 in Eq. 18. See also Jenkins (1987).
For example, although all of the sons and daughters were potential beneficiaries of the guaranteed student loan scheme introduced in 1959 and explicitly subsidized after 1969, only the younger cohorts tended to benefit from the massive expansion in this system after the mid-1980s. Moreover, the various cohorts encountered the effects of the deep recession of the early 1990s at different ages (Hämäläinen and Hämäläinen 2005).
Tests, using the t-bar statistics suggested by Im et al. (2003), reject with very high levels of confidence unit roots in the panel earnings data. See Electronic supplementary material—Appendix E.
Pekkala and Lucas (2007) adopt more flexible forms of the cohort interaction and confirm the absence of any clear trend in family income transmission across cohorts born between 1954 and 1970 in Finland, though a downward decline is apparent among earlier cohorts born from 1930 to 1950.
Controls are included, but not tabulated, for the terms \(a_{ct}^{-1}\), A ct , A pτ, and A pτ/a ct from Eq. 26.
The instruments available include the following: the proportions of school-age children in a region living 0–3, 3–5, or more than 5 km from a school in 1959 and in 1969; a dummy variable for whether the child was affected by the conversion to a comprehensive system, which was introduced on a rolling basis throughout the country, roughly from north to south, between 1972 and 1976 (cf. Black et al. 2005, on Norway; on the Finnish reform, see Pekkarinen et al. 2009); a dummy variable for whether the child’s region of residence at age 18 contained a university town and, if not, distance to such a region (see Conneely and Uusitalo 1998). The partial F statistics for the seven available measures, in a first-stage regression including controls for \(a_{ct}^{-1}\) and A ct , are F(7, 31,501) = 3.98 for sons and F(7, 30,227) = 2.81 for daughters.
Mulligan (1999: S197). The portions in brackets are converted into notation of the present paper.
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Acknowledgements
The authors are most grateful to the Academy of Finland for the partial funding of this study under project number 52198. The paper has benefitted considerably from the constructive comments and suggestions of an anonymous referee as well as those of Joshua Angrist, David Autor, Markus Jäntti, Kevin Lang, and Dilip Mookherjee.
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Lucas, R.E.B., Kerr, S.P. Intergenerational income immobility in Finland: contrasting roles for parental earnings and family income. J Popul Econ 26, 1057–1094 (2013). https://doi.org/10.1007/s00148-012-0442-8
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DOI: https://doi.org/10.1007/s00148-012-0442-8