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New evidence on long-term effects of start-up subsidies: matching estimates and their robustness

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Abstract

The German start-up subsidy (SUS) program for the unemployed has recently undergone a major makeover, altering its institutional setup, adding an additional layer of selection and leading to ambiguous predictions of the program’s effectiveness. Using propensity score matching (PSM) as our main empirical approach, we provide estimates of long-term effects of the post-reform subsidy on individual employment prospects and labor market earnings up to 40 months after entering the program. Our results suggest large and persistent long-term effects of the subsidy on employment probabilities and net earned income. These effects are larger than what was estimated for the pre-reform program. Extensive sensitivity analyses within the standard PSM framework reveal that the results are robust to different choices regarding the implementation of the weighting procedure and also with respect to deviations from the conditional independence assumption. As a further assessment of the results’ sensitivity, we go beyond the standard selection-on-observables approach and employ an instrumental variable setup using regional variation in the likelihood of receiving treatment. Here, we exploit the fact that the reform increased the discretionary power of local employment agencies in allocating active labor market policy funds, allowing us to obtain a measure of local preferences for SUS as the program of choice. The results based on this approach give rise to similar estimates. Thus, our results indicating that SUS are still an effective active labor market program after the reform do not appear to be driven by “hidden bias.”

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Fig. 1

Source: OECD (2015), own calculations

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Notes

  1. For an overview of the importance of SUS programs in OECD countries, see Fig. 1.

  2. For example, effect estimates are provided by Tokila (2009) for Finland, Duhautois et al. (2015) for France, Caliendo and Künn (2011) and Wolff et al. (2016) for Germany, O’Leary (1999) for Hungary and Poland, Perry (2006) for New Zealand, Rodríguez-Planas and Jacob (2010) for Romania and Behrenz et al. (2016) for Sweden. An in-depth review of estimated effects and the institutional setup is given by Caliendo (2016).

  3. For a detailed description of the program before and after the 2011 reform for the NSUS in Germany, estimated short-term program effects and a discussion of the importance of the institutional setup of the program, see Bellmann et al. (2018).

  4. It is currently the only SUS program available to unemployment benefits I recipients. Unemployment benefits II recipients, which are mostly long-term unemployed or individuals with very sparse employment history, are eligible for a different program called “Einstiegsgeld,” which is not the focus of this study.

  5. For a description and evaluations of the predecessor programs, see Caliendo and Künn (2011, 2014, 2015), Caliendo et al. (2016).

  6. Participants spent on average 2.8 months in unemployment before entering the program. Our sample of non-participants was unemployed for 2.7 months on average prior to the assigned date of entry. The p value of a t-test of equality of means is about 0.22.

  7. According to the FEA, about 7400 individuals entered the program between February and June 2012.

  8. The significant gap between treated and comparison group characteristics is due to the fact that pre-matching was done in a very coarse way to ensure minimal overlap between the two groups.

  9. The matching is performed using the psmatch2 ado-package by Leuven and Sianesi (2003).

  10. In the spirit of Imai et al. (2008), a grid search is performed, choosing the bandwidth that maximizes balance by minimizing the pseudo-\(\hbox {R}^2\) after matching. We found this to be the case for \(h=0.13\).

  11. The entire estimation procedure is repeated for each subsample. Balancing indicators and propensity score distributions for the subsamples are available upon request from the authors. Generally, matching quality is somewhat worse due to smaller sample but still within the recommended range of 3-5% in terms of mean standardized bias as given by Caliendo and Kopeinig (2008).

  12. The interval derived by Crump et al. (2009) is optimal in the sense that it minimizes the asymptotic variance of matching estimators. Choosing \(\alpha \) involves a trade-off: Larger \(\alpha \) reduces imbalance and extrapolation leading to lower variance, while discarding information increases variance. As software implementation, we use their accompanying optselect package to obtain \(\alpha \).

  13. The radius matching with bias adjustment is implemented using the radiusmatch package of Huber et al. (2015).

  14. Interestingly, their results suggest a lesser role of personality traits for selection into SUS, which may indeed indicate more severe selection into treatment through caseworkers after the reform.

  15. One additional finding of Chabé-Ferret (2015) is that it is advisable not to condition on pre-treatment characteristics in the matching process when using CDID. However, for our application, this does not make any significant difference.

  16. For binary outcomes, we use the mhbounds package by Becker and Caliendo (2007), and for continuous outcomes, rbounds is employed as described by DiPrete and Gangl (2004).

  17. There are several reasons why we are constrained to contemporaneous data for the instrument. First, data from the previous year correspond to the pre-reform program and thus measure the preference for a nonexistent program. Second, data from the month of January 2012 (the first month after the reform took place) cannot be used as the number of approved applications is contaminated by applicants from before the reform. Third, data after our sampling time frame cannot be used as there was a reform of LEA districts, which led to the disappearance of 22 LEAs. The data on applications for the program and actual entries are obtained from administrative data from the FEA.

  18. The only difference is that we drop interaction terms as these were only included to further improve balance in X across treatment groups D. This choice does not affect our IV results in any significant manner. Results with the interaction terms included can be obtained from the authors on request.

  19. Coefficients on personality traits and test results from the auxillary regression are shown in Table A.3.

  20. The median corresponds to roughly a 50% leave-one-month-out approval rate.

  21. To see this, compare the square brackets in Table 6 with the baseline results from Table 4.

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Correspondence to Marco Caliendo.

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The authors thank Lutz Bellmann, two anonymous reviewers, the editor and participants at the 7th ifo Dresden Workshop on Labour Economics and Social Policy, the University of Barcelona’s Workshop on Unemployment and Labor Market Policies, the 2017 conference of the European Society for Population Economics, the LISER workshop on Causal Inference, Program Evaluation, and External Validity, and the 2017 conference of the European Association of Labor Economists for helpful discussions and valuable comments. We are grateful to the Institute for Employment Research (IAB) for cooperation and institutional support within the research Project No. 1755.

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Caliendo, M., Tübbicke, S. New evidence on long-term effects of start-up subsidies: matching estimates and their robustness. Empir Econ 59, 1605–1631 (2020). https://doi.org/10.1007/s00181-019-01701-9

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