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Demographic change and the labour share of income


Despite similar levels of per capita income, education and technology, the development of labour income shares in OECD countries has displayed different patterns since 1960. The paper examines the role of demography in this regard. We first use a standard overlapping generations model to derive the mechanisms by which demographic change can affect the labour share. It turns out that demographic change can affect the labour share either by altering the domestic capital intensity, by causing factor-biased technological change or in a small open economy framework by creating a gap between domestic savings and investments. The latter affects the country’s investments abroad and in return its net foreign asset income which directly leads to changes in the labour share. Empirical estimations based on these insights, provide evidence that an increases in the expected retirement durations and old-age dependency ratios as well as declines in labour force growth rates have indeed been major forces behind the decline in labour shares that took place in many countries. These effects tend to be larger in open economies and pension reforms towards a funded pension system seem to have accelerated the effects.

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  1. In this paper, we use a definition of the labour share that also accounts for self-employment income, assuming that other categories of workers earn the same average wage as employees. A more detailed description is provided in Section 3.

  2. In reality, virtually all OECD countries operate pension systems that are at least partly PAYGO financed. However, including an ideal-type fully funded pension system as an alternative is the simplest way to investigate what effect on the labour share a pension reform would entail, in which the PAYGO financed component of retirement income is reduced.

  3. Simulations of a calibrated version of this model can be found in Schmidt and Vosen (2010).

  4. Figure 4 exemplifies that current account balances in countries less affected by demographic change tend to be more negative. The effect of demographic change on saving and investment behaviour as well as international capital flows has been subject to a large number of theoretical and empirical studies. By and large all of these studies find that differences in demographic trends have been important determinants of international capital flows. E.g. Higgins (1998), Hebbertson and Zoega (1999) and Lane and Milesi-Ferretti (2001) all examined the links between national age distributions, saving and investment rates, using panel data for a large number of countries. It turns out that countries with a high share of young people tend to run current account deficits, joining the ranks of capital exporters as they grow older.

  5. See Obstfeld and Rogoff (1996). This model is still the main approach for analyzing the economic effects of demographic change (Weil 2008). Since demographic change affects the economy mostly in the medium- to long-run, it seems reasonable to employ a neo-classical model.

  6. The model neglects other potential links between demographic developments and the labour share. For example a growing number of studies analyses the effects of demographic change on economic growth in models with endogenous productivity (Prettner and Prskawetz 2010). However, it is still an open question which approach is best suited to model this link.

  7. In reality, the PAYGO budgets are indeed often unbalanced, i.e. current pensions at least temporarily exceed current contributions. The government can either compensate for the deficit by transferring sources from the overall budget, or by increasing τ. Since there are practical limits to raising τ, this has caused many governments, confronted with rising old-age dependency ratios and thus prospective deficits within the PAYGO pension system, to keep τ constant and shift towards a more funded pension system.

  8. Demographic effects on interest rates are discussed in the literature evolving around the so-called asset meltdown hypothesis (e.g. Abel 2001, 2003; Poterba 2001; Brooks 2002).

  9. Empirical estimates of σ display mixed results. Rowthorn (1996) estimates σ to be substantially lower than one in all OECD countries. Blanchard (1997) estimates σ to be slightly greater than one in Continental European Countries and slightly lower than one in Anglo-Saxon Countries. Berthold et al. (2002) estimate σ to be substantially greater than one in Germany and France and a Cobb–Douglas-like production structure in the US. See also Chirinko (2008) for an overview of empirical studies that estimate σ.

  10. Accordingly a possible “asset meltdown” problem is alleviated to the extent that capital is internationally mobile (Börsch-Supan et al. 2006).

  11. Several methods we use, require the panel to be balanced, so that a number of countries, for which complete time-series of all variables are not available, had to be dropped from the sample. The final sample includes ten countries, namely: Australia, Austria, Belgium, France, Germany, Japan, Netherlands, Norway, Sweden and USA. In regressions with the PAYGO variable we also had to drop Belgium because the variable remains constant over the whole period.

  12. See Breitung and Pesaran (2005) for an overview.

  13. For example the Australian Social Security Act of 1991 entailed changes in retirement age for women born after July 1935 thus applying to those cohorts of women at age 40 since 1975. However, we do not adjust the retirement age for women at age 40 before 1991 when the reform was enacted.

  14. Projected annual data on old-age dependency ratios until 2050 were taken from OECD Social Indicators.

  15. Data were also taken from the AMECO database.

  16. PAYGO schemes are usually defined-benefit programs, but defined contribution schemes are also counted as PAYGO systems, if the social security fund is limited to hold only government debt.

  17. The constant and the coefficients of the first differences are neglected. Equations in which more than one demographic variable was included or in which two interaction terms were included, are not reported in Table 3, as they generally displayed more than one cointegrating relationship. For the same reason we were unfortunately unable to keep the TFP as a control variable in the error-correction models.

  18. If the tests displayed mixed results, we gave the Johansen test priority, accepting those equations as cointegrated in which the absence of cointegrating relationships was rejected at the 1 % level and the presence of less than or equal to one cointegrating relationships was not rejected the at the 5 % level.

  19. It is common practice when reporting cointegrating equations not to indicate significance of the estimated coefficients, as the standard errors are generally biased and their distribution is not asymptotically normal. We nevertheless report the standard errors to get an idea of their sizes relative to the estimated coefficients.


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We thank Christoph M. Schmidt, Christina B. Wilke, Bernhard Mahlberg, Jan Hogrefe, Philipp an de Meulen, Martin Micheli, and two anonymous referees for suggestions and comments.

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Correspondence to Simeon Vosen.

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Schmidt, T., Vosen, S. Demographic change and the labour share of income. J Popul Econ 26, 357–378 (2013).

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  • Labour share
  • Demographic change
  • Panel cointegration
  • Overlapping generations models

JEL Classification

  • E25
  • D91
  • C23