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Honey, I shrunk the kids’ benefits—revisiting intergenerational conflict in OECD countries

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Abstract

Intergenerational conflict may arise when the interests of different age groups do not align. We examine cross-country data to find evidence for this conflict in OECD countries. We derive our results from a FGLS estimation model, which is complemented by a System-GMM estimation to account for potential endogeneity. The data are from a panel of 22 OECD countries over the time period 1985–2005. We find little support for intergenerational conflict in general; however, those who are close to statutory retirement age dislike public expenditure on families and education because, once they retire, they will have less income compared to their work income. This effect is transitory, however, implying a change in voting behavior during retirement age.

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Notes

  1. An expanding welfare state does not per se imply intergenerational conflict. No such conflict arises when public spending continues to be proportional to changing cohort sizes.

  2. In the remainder of the paper, we disregard this argument for empirical reasons as in our setting the national political equilibrium cannot be directly inferred from it.

  3. Unfortunately, the Hansen test is not robust to the number of instruments included. As the number of instruments increases, the Hansen test also tends to accept H 0 too often (cf. Roodman 2009). However, currently there is no rule for determining the optimal number of instruments. Roodman (2009) argues that one should reduce the number dramatically to investigate possible distortions. In the present study, there is a natural upper bound based on the number of countries in the sample. Roodman proposed two possibilities for decreasing the instrument count. The first is to use only some lags as instruments. Hence, we used only lags one and two because the correlation should decrease with the time lag. The second alternative is to ‘collapse’ the instrument matrix. This option creates only one instrument for each not strictly exogenous variable and lag distance. Otherwise, the instrument matrix contains instruments for each variable, lag distance, and time period.

  4. Hence, strictly speaking, the variable is not ‘(log) family benefits per child’ but ‘(log) family benefits per 1,000 children’.

  5. These countries are Australia, Austria, Belgium, Canada, Denmark, Finland, France, Greece, Ireland, Italy, Japan, Luxembourg, the Netherlands, Norway, Portugal, Spain, Sweden, Switzerland, Turkey, the United Kingdom, and the United States.

  6. These countries are Belgium (1995), Luxembourg (1995, 2005), and the United States (1995). We omit these three countries in the subsequent analysis.

  7. Note that we do not have the problem of perfect multicollinearity here because we omit the youngest age group. Furthermore, the two variables Child and SUB524 are not defined with the same denominator. Thus, our control variables do not control for the entire age structure.

  8. The Federalism Index is not available for Turkey. EHII data are missing for Switzerland, the year 2005 for all countries and for Belgium (2000), Luxembourg (2000), and Portugal (1995, 2000). Decommodification cannot be analyzed for Greece, Luxembourg, Portugal, Spain, or Turkey because of missing observations.

  9. We use V70up here to allow for a better comparison with the remaining regressions in Table 3. However, we also tested for the effect of the group of all retirees (i.e., V65up) and found analogous insignificant coefficients, with the exception of family benefits as a percentage of GDP, where the coefficient of V65up is weakly significant and negative (this is due to the inclusion of those aged 65–69, who strongly oppose family benefits, as shown below).

  10. The negative (but insignificant) coefficient in the ‘per child’ specifications (regressions (4)–(6)) may be interpreted as a reflection of Becker’s (1960) quantity-quality tradeoff with respect to children. Under the ceteris paribus condition, the same share of GDP is assumed to be allocated to family benefits, i.e., total spending is said to remain unchanged while the population share of children is rising. Hence, more children have to be accommodated by the same amount of resources, implying that spending per child must fall.

  11. This is in contrast to Brunner and Balsdon (2004) and Cattaneo and Wolter (2009), who argue that retirees’ remaining life expectancy is probably too short to selfishly support investments that increase the pie. However, these studies are not directly comparable to our approach as they use survey data at a sub-national level in single countries (the United States, Switzerland). Our findings are thus closer to Lindert (1996) and Busemeyer (2007).

  12. GDP per capita might be viewed as a truly endogenous variable. Taking account of this in our System-GMM estimation does not change our results, however, so we remain with our initial estimation model. The results of this exercise are available from the authors upon request.

  13. The F-statistic for each regression is reported to test for the overall significance of the variables. However, the Hansen test is not the χ 2 test statistic but the p-value of the corresponding test. The reason for this change in notation is that a large p-value, i.e., a value close to one, indicates a misspecification of the test. This is because the Hansen test is not robust to the inclusion of too many instruments (Roodman 2009). We can also infer directly from the p-value that the test is never significant at conventional levels. Thus, we can accept the null hypothesis of exogeneity of the instruments.

  14. Note, however, that we do not obtain this result for the ‘per child’ specification (not shown here).

  15. The per child and per school-ager one-way fixed effects regressions, which are not shown here, reveal that the effect of the V6569 age group is more dominant on educational expenditure than on family benefits. However, the results are not so different that our previous conclusions change.

  16. Lott and Kenny (1999), for the United States, and Aidt and Dallal (2008), for Western Europe, show substantial increases of the social spending share in GDP as a consequence of women’s suffrage.

  17. The precise drop in consumption around the time of retirement is estimated to be substantial. Mariger (1987) estimates a reduction of 43 % and Bernheim et al. (2001) of 10–20 % for the United States. Banks et al. (1998) finds a 35 % decline for the United Kingdom.

  18. Our argument should not be misunderstood as a case for welfare state expansion due to presumed old-age poverty. Retirees’ wealth may be sufficiently high to guarantee a decent living; nevertheless, any drop in replacement income may be viewed as an unpleasant reduction in the standard of living. This appears to be especially true when children have not yet ‘flown the nest’.

  19. To provide a first impression of what we might expect, we included Polity IV’s (Marshall and Jaggers 2008) democracy variable into our empirical model. Obviously, our country sample does not show a lot of variation in this variable, so it does not come as a surprise that the results of this exercise are not significantly different from our benchmark estimations. However, at least education expenditure per GDP seems to be significantly negatively affected by the democracy variable in some specifications, indicating that intergenerational conflict only arises when the elderly have the opportunity to out-vote the young. All results are available from the authors upon request.

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Acknowledgements

The authors are indebted to two anonymous reviewers and the editor, William F. Shughart II, for their excellent comments.

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  1. Department of Economics, University of Freiburg, Wilhelmstr. 1b, 79085, Freiburg i. Br., Germany

    Tim Krieger

  2. Ifo Institute—Leibniz-Institute for Economic Research, University of Munich, Poschingerstr. 5, 81679, Munich, Germany

    Jens Ruhose

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Krieger, T., Ruhose, J. Honey, I shrunk the kids’ benefits—revisiting intergenerational conflict in OECD countries. Public Choice 157, 115–143 (2013). https://doi.org/10.1007/s11127-013-0064-z

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