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Asymmetric information in external versus internal promotions

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Abstract

Employers have an incentive to fill managerial positions with the most able candidates, and they face a central choice of promoting from inside or outside the firm. With respect to the observability of the candidates’ quality, individual and job-specific characteristics might exert differential influences on these two channels of promotion. We conduct a close comparison of co-workers within occupation–workplace cells in multinomial logit regressions and thereby assess the drivers of external and internal promotions. The results demonstrate that overtime hours, which are observed only by the current employer, are more important for internal promotions, whereas formal educational degrees are more relevant for promotions across establishments. Moreover, the promotion gap for women and foreigners is large and particularly pronounced for promotions across workplaces. However, the disadvantage of women decreases when they work overtime hours.

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Notes

  1. There are numerous ways for a manager to obtain information about potential candidates for a promotion to a managerial position. For internal promotions, managers usually have more sources of information. For instance, they learn about actual hours worked from direct observation and/or by means of time recording, if available. For promotions from the outside, managers have to rely on other resources, e.g., CVs, interviews, or information deliberately made public by a person via a professional networking platform.

  2. In our analysis, only the part of actual working hours that represents a discretionary choice of the worker is relevant. In this way, by controlling for establishment heterogeneity, we try to purge our estimation results from the impact of employer-induced changes in actual working hours.

  3. Only covariates capturing heterogeneities of the destination job are measured in period t + 1.

  4. Wooldridge (2010) suggests the use of so-called Mundlak terms, i.e., group-specific averages, to capture correlated unobserved heterogeneity in nonlinear model. This procedure approximates the same identification strategy as fixed-effect estimation or regressing on group-specific dummy variables, which is infeasible in large-scale nonlinear models such as the presented multinomial logit.

  5. We choose being non-promoted at the same workplace as the base outcome because in this case, no change is observed for the individual.

  6. We observe the population of employed individuals liable to social security contributions; thus, we exclude individuals in the black market, the self-employed, and civil servants. However, promotions might be rather limited on the black market, and they are irrelevant for self-employment.

  7. Moreover, since top manager are often employed under a contract for work which does not constitute employment subject to social security, presumably only a non-random subsample of top-managers is included in our administrative data.

  8. Due to our data sampling approach, we observe the entirety of promotions but only a random sample of non-promotions. Consequently, the shares of promotions among all observations are artificially high and do not refer to the true proportions in the population.

  9. Suppose that for a specific characteristic z = 1, we are interested in the relative risk of being promoted (y = j) in comparison with staying in the base category (y = base), where category y = j is overrepresented by the factor \( F_{j} \left( {z,x} \right) \),

    i.e., \( {\text{rrr}}^{z = 1} (y_{i} = j | x_{i} ) = \frac{{\frac{{Pr\left( {y = j |z_{i} = 1,x_{i} } \right)*F_{j} \left( {z,x} \right)}}{{Pr\left( {y = {\text{base}} |z_{i} = 1,x_{i} } \right)}}}}{{\frac{{Pr\left( {y = j |z_{i} = 0,x_{i} } \right)*F_{j} \left( {z,x} \right)}}{{\Pr \left( {y = {\text{base}} |z_{i} = 0,x_{i} } \right)}}}} \). As long as the overrepresentation is not structurally correlated with the covariates, i.e., Fj(z,x) = Fj, the baseline probability of each outcome category is canceled out of the relative risk ratio and does not influence the promotion probabilities of interest.

  10. While most of the theoretical considerations build on a firm-level argumentation, we use an establishment-level distinction in the data. If the hypothesized information is available across establishments but not across firms, the establishment-level distinction might be rather imprecise, leading to an underestimation of the difference between internal and external promotions.

  11. A more comprehensive but after these revisions slightly outdated overview on the administrative employment register and its collection process is provided in Oberschachtsiek et al. (2009).

  12. All our baseline results can be replicated using a reduced 2% random sample of non-promoted workers from within the same workplaces, ensuring that the sampling does not affect the presented results.

  13. For a robustness check, we utilize a 2% random sample of all non-promoted employees of the entire work force, i.e., employees working at any establishment during that time. While this subsample does not allow us to control for establishment heterogeneity, the results remain unchanged from those in panel A of Table 2.

  14. Since we use a cross section, the detailed controls for age also capture cohort effects, which are relevant for promotions (Kwon et al. 2010).

  15. The churning rate is defined as the employment neutral turnover rate at the establishment level.

  16. The number of promoted individuals is calculated by taking the shares of promoted individuals 0.074 (internally promoted) + 0.184 (externally promoted) times the total number of individuals 386,048.

  17. Both of the presented specifications control for observable establishment characteristics, such as establishment size, industry, and churning rate, as well as the number of managers and promotions, at both the initial and destination establishments. Hence, the observables control for differences in hierarchical levels and job turnover in managerial positions.

  18. The likelihood ratio test provides a possibility to compare both models. The test statistic is 7786, which is evaluated at a Chi-squared distribution with 384 degrees of freedom since the full model estimates 384 additional coefficients. Hence, the equivalence of both models is clearly rejected.

  19. See also Appendix A for the difference in actual working hours across and within workplaces.

  20. In Table B1 of Appendix B, we can show that the differential effect on external versus internal promotions is fully robust to the use of secondary schooling degrees instead of tertiary vocational certificates.

  21. An alternative theoretical explanation suggesting the same outcome is provided by a large stream of theoretical contributions addressing promotions as a signal (e.g., DeVaro and Waldman 2012; Cassidy et al. 2016; Bernhardt 1995), where employers have an incentive to promote educated workers before their uneducated but equally or even more capable counterparts.

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Authors and Affiliations

  1. Institute for Employment Research (IAB), Regensburger Str. 100, 90478, Nuremberg, Germany

    Mario Bossler & Philipp Grunau

  2. Labor and Socio-Economic Research Center (LASER), Nuremberg, Germany

    Mario Bossler

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  1. Mario Bossler
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Correspondence to Mario Bossler.

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We gratefully acknowledge helpful comments of two anonymous reviewers and the editor. Moreover, we thank Silke Anger, Lutz Bellmann, Christian Pfeifer, and Claus Schnabel as well as participants at the seminars in Nottingham and Nuremberg, the 2015 Colloquium on Personnel Economics, the 2015 Annual Conference of the Scottish Economic Society, and the 2015 Spring Meeting of Young Economists for helpful comments and suggestions. We acknowledge valuable assistance in data provision by the IT Management Department of the Institute for Employment Research. The usual disclaimer applies.

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Bossler, M., Grunau, P. Asymmetric information in external versus internal promotions. Empir Econ 59, 2977–2998 (2020). https://doi.org/10.1007/s00181-019-01749-7

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