Abstract
A major concern regarding the consequences of offshoring is the worsening of the labour market position of low-skilled workers. This paper addresses this issue by providing evidence on the impact of offshoring on the skill structure of manufacturing employment in Belgium between 1995 and 2007. Offshoring is found to significantly lower the employment share of low-skilled workers. Its contribution to the fall in the employment share of low-skilled workers amounts to 35 %. This is mainly driven by offshoring to Central and Eastern European countries. While most of the previous papers on this subject focus on materials offshoring, we show that offshoring of business services also contributes significantly to the fall in the low-skilled employment share. As a complement to the existing literature, we compare the widely used current price measure of offshoring with a constant price measure that is based on a deflation with separate price indices for domestic output and imports. This reveals that the former underestimate the extent of offshoring and its impact on low-skilled employment. Finally, we also find that the impact of offshoring on low-skilled employment is significantly smaller in industries with a higher ICT capital intensity.
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Notes
In theoretical contributions, the assumption of labour market clearing ensures full employment through adjustments in relative wages (Baldwin and Robert-Nicoud 2010). In the empirical contributions surveyed below, identification is generally achieved through an exogenous wage rate.
For broad offshoring, all imports of intermediates are taken into account for each industry, while for narrow offshoring, only intermediates from the same industry are considered.
It is, however, not entirely clear in this paper whether the authors replicate the standard offshoring measure of Feenstra and Hanson (1996) or just use total imports for the products corresponding to the industries in the sample.
For the sake of brevity, the industry list is not given here but can be obtained from the authors upon request.
Some authors divide by output, e.g. Ekholm and Hakkala (2006), and others by value added, e.g. Hijzen et al. (2005). The main weakness of this indirect measure is that it does not take into account cases where an activity is shifted abroad without subsequently giving rise to imports and cases where the final stage of the production process is shifted abroad.
Standard Classification of Products by Activity in the European Community (CPA2002 version).
The data on the geographic distribution of imports come from Intrastat and Extrastat for goods and from the balance of payments for services.
Austria, Australia, Canada, Denmark, Finland, France, Germany, Greece, Iceland, Ireland, Italy, Japan, Luxemburg, the Netherlands, New Zealand, Norway, Portugal, Spain, Sweden, Switzerland, the UK and the US. These countries plus Turkey were the OECD member states by the middle of the 1970s.
Bulgaria, the Czech Republic, Estonia, Hungary, Latvia, Lithuania, Poland, Romania, the Slovak Republic and Slovenia.
China, Hong Kong, India, Indonesia, Malaysia, the Philippines, Singapore, South Korea, Thailand and Taiwan.
This implies efficiency gains, notably because the number of parameters to be estimated is lower.
Hours worked by skill level are not available for Belgium. Hence, just like all other paper in this literature, we use data on the number of persons.
We refrain from including variables that may indirectly also account for technological progress, e.g. a time trend.
According to a Hausman test with clustered standard errors, a random effects model would not be appropriate.
Using fixed effects estimation with a panel where the number of industries N largely exceeds the number of time periods T implies that the coefficients of our equation should be interpreted as structural elasticities rather than short-run or long-run elasticities given that the estimation essentially relies on the variation between industries for the variables.
As an alternative, we have also tested the instrumental variable approach suggested in Canals (2006), which relies on contemporaneous and lagged weighted averages of the relative wage for all other industries as instrument for an industry’s relative wage. The results are similar to those below, but instrument validity is rejected. We do not report these results here, but they are available upon request.
It is in fact an exogeneity test, i.e. “under the null hypothesis the specified endogenous regressor can actually be treated as exogenous” (Baum et al. 2007, p. 482). The test reported in Table 4 is equivalent to a C or GMM distance test where the test statistic is distributed as a \(\chi ^{2}\) with a number of degrees of freedom equal to the number of potentially endogenous regressors, and, with homoskedastic errors, it is identical to the Wu–Hausman \(F\) test for endogeneity (Baum et al. 2003, pp. 23–25). In our case, it is necessary to account for clustered standard errors due to the higher level of aggregation of the R&D intensity variable.
The R&D-intensity is defined as the industry-level R&D stock divided by output. Its level of aggregation is 2-digit Nace Rev.1.1 (16 industries for manufacturing) instead of the more detailed classification with 63 industries.
As explained in Baum et al. (2003, p. 11), estimation by gmm is generally more efficient than 2sls estimation due to the use of the optimal weighting matrix. However, the estimation of this matrix requires a large sample size and the properties of the gmm estimator may therefore be poor in small samples, notably leading to over-rejection of the null hypothesis in Wald tests.
The p value of the R&D intensity amounts to 0.11 in column (a). Without the cluster correction, it would be significant at the 5 %-level. However, it should be noted that in our case the R&D intensity would contribute to raising the low-skilled employment share given the overall fall in the R&D intensity in manufacturing between 1995 and 2007 (see Table 10 in the Appendix). The contribution of the R&D intensity would, however, be small (\(<\)0.5 %).
The two variables are not jointly significant either: the p value of a joint Wald test for the R&D intensity and the ICT capital stock is 0.1719.
Own-price elasticities for high-skilled and low-skilled labour for these regressions can be found in columns (b)–(d) of Table 12 in the Appendix. They are very close in terms of size to those for the standard specification in column (a). To complete the results, we have also run gmm estimations for the specifications in columns (b)–(d) of Table 6 (see columns (b)–(d) in Table 11 in the Appendix). There are no substantial differences compared with the 2sls estimations.
We have also interacted the R&D intensity with the high-tech dummy, but this did not produce significant results.
Wald test for joint significance of: OM and OM * Hitech: test-stat \([\chi ^{2}(1)] = 10.67\), p value \(\,=\,0.004\); OS and OS * Hitech: test-stat [\(\chi ^{2}(1)] = 34.51\), p value\(\,=\,0.005\).
In low-tech industries, the contribution to the fall in the employment share of low-skilled workers is close to 5 % for the average increase in materials offshoring in these industries.
Due to the difference in the average increase in business services offshoring, the contribution to the fall in the low-skilled employment share amounts to approximately 10 % for both.
For both materials and business services offshoring, the coefficients of the offshoring variable and the respective interaction term with the ICT capital intensity are jointly significant. Wald test for joint significance of: OM and OM* ICT_VA: test-stat \([\chi ^{2}(1)] = 20.62\), p value\(\,=\,0.000\); OS and OS * ICT_VA: test-stat [\(\chi ^{2}(1)] = 44.04\), p value\(\,=\,0.000\).
For the average ICT capital intensity, the contributions to the fall in the low-skilled employment share—computed based on the average increase in materials and business services offshoring—are in line with the standard specification: 2 % for materials offshoring, and 10 % for business services offshoring.
We have included offshoring intensities for the three above-mentioned regions as well as the rest of the world (OTHER) in the equation. Moreover, we have decided not to split business services offshoring by region since it is almost entirely limited to the OECD region.
The own-price elasticities for low-skilled and high-skilled labour are very close to those reported in Table 12: respectively \(-\)1.387 (0.167) and \(-\)0.697 (0.0767).
Hence, returns to scale of the dual production function are not constrained (see Berndt 1991, pp. 469–470).
For ease of presentation, time and industry subscripts have been omitted.
Then, given that the elasticities are nonlinear functions of the estimated parameters, the standard errors of the elasticities must be computed by the ‘delta method’.
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Appendix
Appendix
Translog production and cost functions were introduced in the first half of the seventies (Christensen et al. 1971) and have been frequently used in empirical work since then. They allow substitution elasticities to be unrestricted and they are nonhomothetic, i.e. cost-minimizing relative input demands may depend on the level of output (Tables 9, 10).Footnote 34
Denoting total variable costs \(C\), the prices of \(N\) variable input factors \(P_{j}\) and output \(Y\), the general formulation of the translog cost function is as follows:Footnote 35
In a classic KLEMS framework, Eq. (6) represents a five-factor model (\(N=5\)) with capital (K), labour (L) and energy (E), materials (M) and services (S) inputs as variable factors. Labour can further be divided into different skill levels (Tables 11, 12 and 13).
In Eq. (6), \(N(N-1)/2\) symmetry conditions (\(\delta _{jk} =\delta _{kj} )\) can be imposed without loss of generality. Moreover, a ‘well-behaved’ cost function should be homogeneous of degree 1 in prices, i.e. a proportional increase in all variable input prices shifts total variable costs by the same proportion. This implies the following restrictions:
According to Shephard’s lemma, the cost-minimizing input quantities \(X_j \) can be derived by differentiating total costs with respect to the prices of the input factors:
From this, one obtains a set of \(N\) cost share equations, which add up to 1.
The own price elasticities \(\varepsilon _{jj}\) and cross price elasticities \(\varepsilon _{jk}\) are given by:
These elasticities are not constant, but differ at every data point. It is common practice to compute them either for the mean or the first, central or last year of the sample. Estimates of these elasticities should use fitted rather than observed cost shares.Footnote 36
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Hertveldt, B., Michel, B. Offshoring and the Skill Structure of Labour Demand in Belgium. De Economist 161, 399–420 (2013). https://doi.org/10.1007/s10645-013-9218-0
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DOI: https://doi.org/10.1007/s10645-013-9218-0
Keywords
- Materials and business services offshoring
- Constant prices
- Labour demand
- Skill upgrading
- ICT capital intensity