Abstract
This chapter presents a new supply side framework that quantifies the impact of structural reforms on per capita income in OECD countries. It presents the overall macroeconomic impacts of reforms by aggregating over the effects on physical capital, employment and productivity through a production function. It is found that product market regulation has the largest overall single policy impact 5 years after the reforms. But the combined impact of all labour market policies is considerably larger than that of product market regulation. It is also shown that policy impacts can differ at different horizons.
The chapter benefitted from useful comments and suggestions from Andrea Bassanini, Gilbert Cette, Alain de Serres, Sean Dougherty, Falilou Fall, Andrea Garnero, Alexander Hijzen, Catherine L. Mann, Fabrice Murtin and Jean-Luc Schneider. A short version was published in OECD Economic Studies 2016(1), 91–108 and a longer version appeared as OECD Economics Department working paper No. 1354. The views expressed in the paper are those of the author and do not necessarily reflect the opinions of the OECD or any other institution the authors are affiliated with. The underlying work to this chapter has also been published in the OECD Economic Journal, 4(1), 91–108, 2016, under the title “The quantification of structural reforms in OECD countries: A new framework”.
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- 1.
For instance, policy effects on labour market outcomes are analysed for specific policies (independently of other possibly relevant policies) on substantially different country samples. The effects of unemployment benefits, tax wedge and active labour market policies are taken from Bassanini and Duval (2006). The effect of childcare spending reported in Jaumotte (2003) is used. The first study covers 20 countries and the period 1982–2003, and uses OLS and SUR for estimation purposes. The second paper looks at 17 countries and 1985–1999. It employs 2-stage least squares to estimate policy effects. For a more detailed comparison, see Tables A1.1 and A1.2. in Égert and Gal (2016).
- 2.
Appendix 1 in Égert and Gal (2016) provides a detailed comparison of the old and new frameworks.
- 3.
- 4.
The MFP and ETCR series have common trends captured by year fixed effects. These trends are strongly correlated with each other. The correlation between the time fixed effects of MFP and the demeaned overall ETCR series is 0.72 (the series are also demeaned in the regressions including country fixed effects). When we compare the time fixed effects in the MFP and ETCR series, the correlation is 0.77. This is not surprising as time fixed effects explain about 89% of the variation of the demeaned overall ETCR series. When decomposing the overall ETCR indicator into (i) barriers to entry and (ii) public ownership, public ownership survives the inclusion of year fixed effects. This variable could potentially be used for the purpose of quantification (at the expense of covering fewer policy areas).
- 5.
The finding that EPL is statistically not significant stands somewhat in contrast with the literature using sector- and firm-level data relying on difference-in-difference approaches. For instance, Bassanini et al. (2009) finds for a set of 16 OECD countries from 1982 to 2003 that country-level EPL is associated with lower MFP growth in sectors with higher layoff rates. Rincon-Aznar and Siebert (2012) show the negative relation to hold for manufacturing sectors but not for the services sectors. Using firm-level data for the USA, Autor et al. (2007) report mixed evidence on the negative relation between employment protection and the level of MFP: the coefficient estimates are negative but only one coefficient in two is precisely estimated. Dougherty et al. (2011) show that state-level employment regulation lowers MFP levels in Indian firms operating in more-labour intensive industries.
- 6.
This negative relationship is robust to alternative country coverages (for narrower samples composed of more developed OECD countries) and to the definition of the capital stock (real capital stock, capital stock/output, capital stock/workers). This result needs qualification. The effect of EPL on investment is not clear-cut in the existing body of research. The literature reports no evidence that labour market regulation has any impact of investment at the macroeconomic level and for several OECD countries (Kerdrain et al. 2010). There is mixed evidence on the relation between capital stock and labour market regulation at the firm level. There is evidence for European firms that more stringent EPL reduces investment per worker and capital per worker (Cingano et al. 2010). By contrast, for US firms, research suggests higher firing costs (wrongful discharge exceptions) are linked to higher capital stock and capital-to-labour ratios. But the effect becomes negative when state-specific trends are used. A rise in capital may be related to a correction of an earlier downturn and that the introduction of more stringent firing regulations followed a rise of the capital-to-labour ratio (Autor et al. 2007). For Italian firms, estimation results show that the introduction of unjust-dismissal costs raises the capital-to-labour ratio in firms with less than 15 employees, compared to larger firms (Cingano et al. 2015).
- 7.
These results are robust to various sensitivity checks, including different estimation methodologies, control variables and a time period covering only the pre-financial crisis period (Gal and Theising 2015). Nevertheless, jack-knifing the sample, i.e. dropping one country at a time from the sample, shows some sensitivity to the country coverage.
- 8.
The magnitude of the estimated impact (−0.3 for the youth) seems consistent with studies showing elasticities of −0.1 to −0.2 (see recent surveys by Neumark 2015 and OECD 2015b). This is because we use the Kaitz index (median to minimum wage), which in our sample averages at 50% (Gal and Theising 2015). Hence a 1% point increase in it translates into a 2% point increase, on average, for the minimum wage level. Therefore, coefficients obtained when using the level of the minimum wage should be multiplied by two to make them comparable with our coefficients.
- 9.
Our coefficient estimates for prime-age women are larger than those reported in the literature using similar datasets (Addison and Ozturk 2012). The differences may be due to different model specification and data coverage. Therefore, care should be taken when using these estimates for quantification.
- 10.
Appendix 5 in Égert and Gal (2016) discusses alternative reform scenarios.
- 11.
ln(1 − 0.9)/ ln (1 − 0.05) ≈ 45 years. The half-life, i.e. the time over which half of the convergence to the new long-run equilibrium happens, can be calculated as ln(1 − 0.5)/ ln (1 − 0.05) ≈ 13.5 years.
- 12.
In addition to the baseline results based on demographic groups, the predicted impacts for the employment rate and the core set of policies are shown for two alternative approaches: (i) results obtained for the overall employment rate; and (ii) results obtained for skill groups.
- 13.
MFP used for the estimations is calculated as follows:
$$ \ln \left({MFP}_t\right)=\ln \left({Y}_t\right)/\left(1-\alpha \right)-\ln \left({L}_t\right)-\ln \Big({CLF}_t-\alpha /\left(1-\alpha \right)\ln {(K)}_t, $$where CLF adjusts labour input for people working but not living in the country or those working abroad for domestic companies α = 0.33, the standard value in the literature and fixed across countries and over time for ensuring comparability in a simple manner.
- 14.
Considering capital intensity, when r is the real interest rate, the capital-output ratio in equilibrium is given by \( \frac{K}{\mathrm{Y}}=\frac{\upalpha}{\mathrm{r}} \). In a more elaborate setting, the real interest rate can be replaced by the user cost of capital, which includes the relative price of investment goods and corporate taxes as further determinants. In addition, excessive regulation can introduce frictions that suppress capital accumulation—a mechanism that can be captured by product and labour market regulation indicators. As for the employment rate, both labour supply and labour demand determinants enter as policy channels in equilibrium (hence no need to include wages or productivity on top of them).
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Appendix: Calculating Total Policy Impacts on Per Capita Income
Appendix: Calculating Total Policy Impacts on Per Capita Income
3.1.1 Theoretical Considerations
In the new framework, similarly to previous frameworks, structural policies affect per capita income through the supply side components. The appropriate aggregation across the components is straightforward in a standard neo-classical model with a Cobb-Douglas aggregate production of the following form:
with h denoting labour-augmenting (Harrod-neutral) technological progress. Note that the empirical construction of the MFP measure that is used for the estimations relies on the formulation in Eq. (3.1). Footnote 13 However, under the assumption of constant returns to scale, Eq. (3.1) can be rewritten in the following way:
where there is a very close link between multi-factor productivity (MFP) and h:MFP = h 1 − α. Introducing per capita measures and after\vadjust{\pagebreak} some rearrangements, per capita income can be expressed as a function of MFP, the capital-output ratio (\( \raisebox{1ex}{$K$}\!\left/ \!\raisebox{-1ex}{$Y$}\right. \)) and the employment rate (\( \raisebox{1ex}{$L$}\!\left/ \!\raisebox{-1ex}{${N}_{wa}$}\right. \)):
where N pop and N wa stand for total population and working age population, respectively.
The advantage of this formulation is that in a standard setting, all components are separable and independent from each other. Specifically, the capital-output ratio does not depend on either productivity or employment, neither is the employment rate influenced by productivity or capital.Footnote 14
For simulating the effects of changes in policies, the above equation will be used in growth rates:
where Δ captures differences over time, which can be interpreted as percentage changes. As mentioned above, MFP in our empirical framework uses the Harrod-neutral specification. Hence Eq. (3.4) can be rewritten as follows:
Similar to the calculation of MFP a standard value for capital elasticity is set in the simulations (α = 0.33). The last term capturing the share of working age population will be assumed to be unchanged over the simulation horizon. Alternatively, demographic projections by the United Nations could be used over the projection horizon (long-term scenarios project of the OECD, see Johansson et al. 2013).
3.1.2 Practical Considerations
MFP and capital deepening are measured in logarithms, while the employment rate is measured in percentage points (between 0 and 100). The simulation framework requires that the reform impacts are expressed in log-points for each supply side component, Percentage point changes in the employment rate are thus transformed into log-points by dividing the changes in the employment rate by the latest observed employment rate for the working age population \( \raisebox{1ex}{$L$}\!\left/ \!\raisebox{-1ex}{${N}_{wa}$}\right. \) (which was 67% in 2013, averaged across all countries in the sample):
Another issue about aggregation is how to obtain the aggregate employment effect from the demographic and skill groups of the population. Policy effects for these groups are aggregated using the groups’ weight in the working age population. For the illustrative simulations presented in this paper, the population structure of the average OECD country is used in the latest available year.
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Égert, B., Gal, P. (2018). The Quantification of the Effects of Structural Reforms in OECD Countries. In: de Haan, J., Parlevliet, J. (eds) Structural Reforms. Springer, Cham. https://doi.org/10.1007/978-3-319-74400-1_3
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