Employment and Wage Adjustment: Insider–Outsider Control in a Polish Privatization Panel Study

Journal of Comparative Economics 30, 251–275 (2002) doi:10.1006/jcec.2002.1775

Employment and Wage Adjustment: Insider–Outsider Control in a Polish Privatization Panel Study1 Atanas Christev CERT, Heriot-Watt University, Edinburgh, EH14 4AS, Scotland, United Kingdom E-mail: [email protected]

and Felix FitzRoy St. Salvator’s College, University of St. Andrews, St. Andrews, FIFE KY16 9AL, Scotland, United Kingdom Received February 11, 2000; revised March 11, 2002 Christev, Atanas, and FitzRoy, Felix—Employment and Wage Adjustment: Insider– Outsider Control in a Polish Privatization Panel Study New survey data for a panel of Polish firms are used to study employment and wage adjustment in state-owned enterprises, and insider- and outsider-controlled privatized firms. In contrast to earlier studies, dynamic panel data estimators allow for endogeneity of observed variables and partial adjustment to shocks. Asymmetric demand and productivity

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We thank Marek Bednarski, Stanislawa Golinowska, Piotr Kurowski, Hans-Georg Petersen, Christoph Sowada, and Hans-Peter Weikard for their contributions to the survey. Financial support by the European Union’s Phare ACE Programme, Grant P96-6227-R, and by the Komitet Badan Naukowych, Grant 02507, for carrying out the survey study is gratefully acknowledged. The paper was conceived while the first author was at the Institute of Public Finance, University of Potsdam, without the support and hospitality of which this work would not have been possible. We are particularly indebted to Marek Bednarski for numerous discusssions on the privatization process in Poland, to Piotr Kurowski for generous help with the data, and to Hans-Georg Petersen for encouragement and support in writing the paper. Michael Funke and Mike Nolan provided valuable comments early on. We are also grateful to Knut Bartels, Christhart Bork, Simeon Djankov, Hartmut Lehmann, Mark Schaffer, and seminar participants at Potsdam and St. Andrews for helpful comments. Earlier versions of the paper were presented at a CEPR/ZEI Bonn workshop on labor markets in transition in Vilnius (April, 2000) and at the Royal Economic Society meetings in St. Andrews (July, 2000). Two anonymous referees and the editor of the journal have been instrumental in improving the manuscript substantially. We thank Jochen Greiner-Mai for prompt and able assistance. All errors are our own. 0147-5967/02 $35.00

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2002 Association for Comparative Economic Studies Published by Elsevier Science (USA) All rights reserved.

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CHRISTEV AND FITZROY shocks have differing effects across ownership categories that are missed by the simpler and more aggregated specifications used in the previous literature. We confirm rent-seeking behavior in insider-controlled firms and find a significant employment growth-wage effect. J. Comp. Econ., June 2002, 30(2), pp. 251–275. CERT, Heriot-Watt University, Edinburgh, EH14 4AS, Scotland, United Kingdom; St. Salvator’s College, University of St. Andrews, St. Andrews, FIFE KY16 9AL, Scotland, United Kingdom. C 2002 Association for Comparative Economic Studies. Published by Elsevier Science (USA). All rights reserved.

Key Words: privatization; insider vs outsider; employment; wages; transition economies. Journal of Economic Literature Classification Numbers: C23, J23, J21, J31, P2, P3.

1. INTRODUCTION An issue of perennial importance in the transition from state to private ownership has been the distinction between insider and outsider control. To ease the political path of transition, some countries, e.g., Russia, have opted mainly for insider privatization. However, most other countries have used a variety of privatization methods, whereby outsiders as well as insiders participated with relative dominance of ownership concentration; see Frydman et al. (1995), Earle and Estrin (1995) for a summary, and Frydman et al. (1999). Some have followed a big-bang approach, e.g., the Czech Republic, others have restructured gradually; while at the same time all have attempted to ensure the emergence of a sound financial sector to facilitate privatization progress and restructuring. In the process, differences in motivation and goals arise between insider and outsider control of privatization; for an evaluation see Blanchard and Aghion (1996) and Blanchard (1996, 1997). These experiences in Central and Eastern Europe provide a unique natural experiment for large scale testing of long-standing hypotheses in industrial organization. A few empirical studies have attempted to assess systematically the effects of ownership control and privatization method on employment and wage adjustment at the enterprise level (Konings et al., 1996; Basu et al., 1997, 2000; K¨ollo, 1998; and Estrin and Svejnar, 1998). The variety of ownership types within most transition countries, and in Poland in particular, offers rich possibilities for understanding the behavior of insider vs outsider firms and for assessing the different outcomes for economic transformation. This paper uses new survey data for a panel of Polish firms during the period from 1994 to 1997 to study employment and wage adjustments and emphasizes explicitly the distinction between insiders and outsiders. In contrast with earlier work on wage behavior and employment in Poland by Commander and Dhar (1998) and Grosfeld and Nivet (1999), we use the stability of ownership varieties achieved or anticipated by the beginning of the period to employ dynamic panel estimation methods systematically. In this way, we correct for the endogeneity of key explanatory variables by using Generalized Method of Moments (GMM) estimators following Arellano and Bond (1991) and Blundell and Bond (1998). These estimators are consistent and considerably more efficient in the context of dynamic panel data models than the others suggested earlier in the literature

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or used in the above studies. Thus, we allow both for the partial adjustment of employment and wages to exogenous shocks, which also reflect the anticipation of privatization, and for the endogeneity of most of the variables by exploiting the full panel rather than using a succession of cross-sectional estimates as in earlier work. Among the new results obtained, we find a significant employment growth-wage effect that seems to have been hitherto neglected in the transition literature. The strong employment growth-wage effect suggests that rapidly growing firms may need to offer premium wages to attract quickly workers from nearby firms or from outside the local labor market. We obtain asymmetric responses to demand shocks2 in growing and declining firms, by various measures, and different responses to demand growth by ownership categories. We confirm rent-seeking behavior in insider-controlled firms as theory suggests and find higher responsiveness to market pressures by outsiders as expected. Thus, we explore systematically the responses of employment and wages in differing ownership categories to shocks and obtain direct evidence of initial labor hoarding. In Section 2, we outline the institutional framework; we introduce the sample and data and discuss our testable hypotheses in Section 3. Section 4 describes the theoretical framework and specification. The estimation and results are presented and discussed in Section 5; Section 6 concludes with a summary of the main findings of the paper. The Appendix provides further details on the firms in the survey. 2. INSIDERS VS OUTSIDERS: THE INSTITUTIONAL SETTING IN POLISH PRIVATIZATION In Poland, ownership transformation of industrial firms through a variety of privatization methods started in 1991. While privatizing the state sector has been relatively slow, the emphasis was placed on commercialization and adequate corporate governance, in contrast to Czech and Hungarian practice.3 Earlier studies (Pinto et al., 1993; Blanchard, 1994; Estrin et al., 1995; Konings et al., 1996; Nuti, 1997; Commander and Dhar, 1998) have documented these changes. However, little has been done to assess how employment and wage responses vary across different forms of ownership and privatization. Privatization is perceived as one of the main tools to create a market-oriented economy; restructuring and change at the enterprise level should be studied in this context.4 Privatization, or the anticipation thereof, leads to effective restructuring of firms only if it is accompanied by the emergence of new owners, i.e., outside investors, with a strong profit motive rather 2 By demand shocks, we mean the real sales or output changes in the product markets experienced by the firms in the intervening period. However, supply shocks are also a possibility. 3 Estrin and Svejnar (1998) provide additional insights and a comparison among countries in the early stages of transition. 4 Djankov and Murrell (2000) provide a comprehensive overview of the empirical literature on enterprise restructuring and its determinants in transition economies.

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CHRISTEV AND FITZROY TABLE 1 Ownership and End-of-Period Employment (in Percent) Enterprise ownership Employment

SOE

NIF

PRII

PRIO

Total

Below 500 employees 501–1000 employees Above 1001 employees

54.2 27.1 18.6

21.2 48.5 30.3

55.8 34.9 9.3

48.3 27.6 24.1

47.0 33.5 19.5

Total

36.0

20.1

26.2

17.7

100

Note. SOE, state-owned enterprises; NIF, firms included in the National Investment Funds; PRII, privatized insider-owned firms; PRIO, privatized outsider-owned firms.

than by employee, or even manager, insiders (Boycko et al., 1996; and Frydman et al., 1999). In our survey, we have data from a panel of firms with diverse forms of ownership that have undergone a variety of privatization methods by 1998. All of these methods influenced the adjustment behavior of employment and wages at the enterprise level by driving a wedge between the motivations of different interest groups within the enterprise in the process, e.g., insiders vs outsiders vs the government.5 However, the delay in privatization of large firms in Poland does not mean that little change has occurred. Market-oriented responses after price and trade liberalization and the withdrawal of subsidies can be observed quite independently of ownership.6 The process of reactive restructuring, i.e., the adjustment of wages and employment, started even before privatization. Pinto et al. (1993) point out that, even before any change in corporate governance has taken place, managers raise their expectations in anticipation of privatization and envisage their performance being rewarded once privatization occurs. An issue of importance in empirical studies, often recognized but neglected in practice, is the problem of endogeneity of ownership (Konings et al., 1996; Estrin et al., 1999). This may arise because the choice of privatization method is not independent of firm performance. It may seem likely that insiders managed to select the better (and smaller) firms that are easier to acquire and control. Table 1 indicates that insiders indeed tended to acquire smaller firms. It is less clear why the firms chosen for the National Investment Funds (NIFs) were not consistently poor performers during this period as we document below. In 5 A growing body of literature has studied these issues in detail, for example Schleifer and Vishny (1994); Aghion et al. (1994); Boycko et al. (1996); and Blanchard and Aghion (1996). For an overview of the Polish experience, see Pinto et al. (1993) and Ipsen and Puntillo (1998). Bornstein (1999) provides a comparative analysis and contrasts with the Czech Republic and Hungary. 6 Aghion et al. (1994) and Aghion and Carlin (1997) provide a detailed discussion. For more on the privatization effects in preprivatization firms, see Estrin et al. (1995). The conclusion of these authors is that long-term adjustment strategies at the enterprise level are closely related to privatization procedures. In this context, see also Roland (1996) on the sequencing of restructuring and privatization.

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this paper, we attempt to understand how and why once privatization, or the anticipation thereof, occurs, it has impacted firm wage and employment behavior taking account of the endogeneity of both. Our estimates suggest that initial conditions matter for the sampled firms. In addition, we test and verify empirically the idea that the anticipation of privatization had an important impact on firm restructuring. The relative diversity of ownership forms in transition economies is evident from our survey, i.e., state-owned firms and privatized firms, and, within this group, we distinguish between employee/manager-controlled and outsider-controlled privatized firms, and new private firms. The newly established private firms appear to have drastically different and more dynamic behavior than those in the rest of the sample. Since we have no new findings to report and only a few observations in this category, we exclude these new private firms from our analysis. In our sample, state-owned enterprises (SOE) include traditional state-owned firms, joint-stock companies wholly owned by the State Treasury, and two firms with remaining majority stakes owned by the state.7,8 The emphasis we put on distinguishing between the types of privatized firms comes from issues raised by a number of earlier studies and is related to the specificity of the ownership transformation process in Poland. Privatized firms that are insider-controlled (PRIIs) include firms that are mainly domestically owned with strong representation of insider interests. Although it is difficult to disaggregate the information obtained from the survey, most of these firms are enterprises that were privatized through liquidation-leasing, a process by which management/employees receive or obtain, often against debt, a controlling share of the enterprise. These are essentially management/employee buyouts. In the sample period, most of these had been privatized and ownership had been consolidated by 1996. The advantage of insider-dominated privatization is the close cooperation between management and employees because of little difference in motivation and goals. However, the inherent disadvantage is that insider interests may lead to increased costs through excessive wage increases, labor hoarding, poor management decisions, and insufficient deep restructuring in the form of new product lines and new investments.9 7

This procedure was used in Poland as a first intermediate step before final privatization; some of these companies were included in mass privatization later and appear in our sample. We have also separated these commercialized firms from SOEs. The results did not yield significant differences. In some cases, managers indicated that they anticipated capital privatization as the discussion on the next page reveals. Thus, we have included the ones with such characteristics in the outsider category for estimation. We also have four joint stock companies with 100% state ownership. When considered as SOEs, the results did not change. We owe this point to an anonymous referee. 8 Since the state remains a passive owner (Frydman et al., 1999) and because of the method anticipated in privatization, we have included these in the privatized outsider firms for estimation purposes. An anonymous referee suggested this treatment. It is important to point out that outsiders changed management much more frequently than insiders. However, about 70% of our insider-dominated firms did not change management at all. 9 Havrylyshyn and McGettigan (1999) and Bornstein (1999) discuss these issues. However, insiderowned firms may not be uniformly poor performers; for a discussion and some evidence on this point, see Carlin et al. (2001).

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Privatized firms that are outsider-controlled (PRIOs) were privatized early in the transition, mostly by 1995 in our sample, using some form of sale to a core group of outside investors. Two distinct procedures of privatization were used to accomplish this goal, namely, direct sales and capital privatization. Some of these firms acquired foreign direct investment in the process of privatization, 9.6% of the enterprises in the sample;10 more than half of these investments have occurred in the period from 1994 to 1997. Outsider involvement is expected to infuse fresh blood into the enterprise by bringing in better management, new knowledge, foreign expertise, and new capital. In due course, wage and employment adjustments, more in line with Western competitors, would be induced. Firms privatized in Poland’s capital privatization program underwent some preprivatization restructuring, as evidenced by managers’ responses in 52% of the commercialized firms in our survey, in the belief that better performing enterprises would be privatized first. The delay in the mass privatization of Polish firms until 1995 provides an exclusive group of enterprises that are owned by 15 government-sponsored National Investment Funds. The objective of the mass privatization program is to transfer a majority of state assets into private hands by equal access vouchers. In the interim, enterprises allocated to this program have been able to restructure, under the direction of the funds, in anticipation of final privatization (Estrin et al., 1999). However, what distinguishes them from the other categories of privatized firms is that they have not yet established a clear governance structure. The major disadvantages with mass privatization are the resulting dispersed structure of ownership claims and the lack of concentrated power in corporate decision making (Havrylyshyn and McGettigan, 1999).11 Whether insiders or outsiders will prevail in the end seems to matter. 3. DATA AND HYPOTHESES We have a panel of 178 firms in Poland for the period from 1994 to 1997. The adjustment of employment and wages at the enterprise level depended crucially on the involvement of firms in the ongoing privatization process. Our sample of firms was selected randomly to investigate patterns of firms’ restructuring in the process of transition and a survey was prepared to address some of these issues. The survey was conducted in early 1998; selected enterprises had to meet two criteria: the number of employees had to be above 300 and the enterprise had to be operating in the period of study. The survey was structured so that a full 10 Only one of our management/employee buyout firms has been privatized with strong direct involvement of foreigners. For the purposes of our analysis, we included this firm in the outsider group. 11 Our reading of the literature, e.g., Pinto et al. (1993), Roland (1994), Blanchard (1994), Ipsen et al. (1998), and Bornstein (1999), suggests that NIFs had little incentive to restructure, given institutional characteristics, which implies poor performance. However, despite these widely noted deficiencies with regard to their corporate governance, the hypothesis is not borne out by our results. One plausible interpretation was suggested to us by an anonymous referee: namely, ownership structure in this category is not more dispersed than it is in other privatized firms.

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TABLE 2 Growth Rates of Sales, Employment, Productivity, and Wages over Entire Period Mean 1994–1997 (in percent) Growth rate

SOE (N = 59)

NIF (N = 33)

PRII (N = 43)

PRIO (N = 29)

Total (N = 164)

Sales Employment Labor productivity Real wages

−2.1 −17.4 15.3 19.7

15.4 −10.4 25.8 24.9

13.7 −6.9 20.6 25.9

3.2 −20.7 23.9 28.3

6.5 −13.8 20.3 23.9

Note. SOE, state-owned enterprises; NIF, firms included in the National Investment Funds; PRII, privatized insider-owned firms; PRIO, privatized outsider-owned firms; N, number of firms in the different categories.

variety of existing ownership forms was included. After a pilot survey to which the initial response was poor, some of the enterprises were dropped and other new ones were selected. The pool of firms meeting the above criteria is the largest possible given resource constraints. Then 370 enterprises were drawn randomly from over 1800 enterprises. In this way, answers were obtained from enterprises of different size and from different industries at various stages of transition and privatization. The sample includes large, mostly manufacturing firms (81.9%) with average employment in the beginning (end) of the period of 753 (664). For the estimation procedure, we use a balanced panel of 164 firms omitting firms with incomplete answers and missing or nonreliable data. We also excluded new private firms from our analysis. Summary statistics for the entire period are presented in Table 2. As in earlier studies,12 we observe certain stylized facts for different categories of ownership; these are corroborated in our empirical results. Over the entire period, we observe an unambiguous rise in real wages, i.e., wages deflated by the producer price index with 1993 as a base year, and a substantial decline in employment. Sales of stateowned enterprises (SOE) declined most and these firms performed worst in the sample. As a result, average real wage increases in SOEs were lower, although these increases were still well above labor productivity gains. Compared with firms in the rest of the sample, only SOEs had decreasing sales, on average, between 1994 and 1997. Distinct differences emerge between firms privatized to insiders (PRIIs) and those privatized to outsiders (PRIOs) in terms of sales and employment growth. Outsider owners reduced employment by more and paid higher wages than any other category. PRIIs had a higher growth rate of sales than those with outside owners. In the beginning of the period, the insider-controlled enterprises increased 12 Pinto et al. (1993), Commander and Dhar (1998), and Grosfeld and Nivet (1999) study Polish firms; Estrin and Svejnar (1998) and Aghion and Carlin (1997) summarize evidence for several other Central and East European countries.

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real wages more than other firms as expected. However, over the entire period, PRIIs have cut employment least, well below the average for the sample, while maintaining slightly above average real wage growth. On the other hand, wage growth has been highest in PRIOs over the period but, for this ownership category, it has also been accompanied by higher than average labor productivity growth. Table 2 also reveals, perhaps unexpectedly, that NIFs outperformed the rest of the sample in both sales growth and labor productivity. As evident, NIFs consistently paid real wage increases below labor productivity gains. It is worth noting that NIFs have also the smallest declines in employment at the beginning of the period and shed labor fastest at the end of the period, implying that adjustment for this type of firms has been significant. This diversity of firms in our sample allows us to test several different hypotheses. First, we characterize an asymmetric-response hypothesis that involves differing elasticities of employment growth with respect to sales growth. As a measure of downsizing and restructuring, this elasticity should indicate a larger significant response to negative demand shocks in PRIOs that tend to respond to market pressures quicker. Earlier studies suggest restructuring is minimal for firms in mass privatization programs (Aghion and Carlin, 1997); thus, we expect little response to sales growth by NIFs. In addition, we identify the effect of privatization and adjustment on the employment response to demand shifts; in turn, this provides evidence on unobservable labor hoarding as the initial response to falling demand. Labor hoarding is defined as the labor that would not have been employed had the firm, or the economy, operated under market conditions at the given level of output. In the context of the transition firm labor hoarding implies a small or insignificant employment growth response to positive demand shocks. In effect, these shocks are accommodated by either labor hoarding, most likely in SOEs, or by increasing employment, most likely in PRIOs. Next, we expect larger employment growth elasticity with respect to wage growth in privatized firms due to the greater pressure to minimize costs. Earlier studies find the expected negative employment-wage elasticity but their lack of attention to endogeneity means that these results must be confirmed. We also expect that strong trade unions should have a moderating influence on employment declines. Hence, we test for the presence of union, firm-size effects and investigate whether larger firms do tend to grow slower, other things equal, as Commander and Dhar (1998) suggest. Another asymmetric response concerns wage growth elasticities with respect to productivity growth, which is a measure of insider power (Grosfeld and Nivet, 1999). Wage growth response is expected to be largest in SOEs and in PRIIs because constraints on rent seeking in these ownership types are weakest. Limitations on wage reduction imposed by alternative employment may imply that the response to negative productivity shocks is likely to be relatively small or insignificant.

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The wage curve hypothesis examines the influence of local labor market conditions on wage outcomes at the firm level. In our study, we expect negative effects of regional unemployment rates on wage growth. While outsiders tend to increase wages commensurate with productivity growth, we also expect higher wages or wage growth for insider-dominated firms. Finally, Western firm-level data establish that larger firms pay higher wages, other things equal. An alternative version suggests that faster employment growth is more likely to push up wages, because new hires must be attracted from existing alternative employment. These firm sizewage effects can be distinguished with both levels and differenced specifications, as we discuss and test in our sample below. 4. THEORETICAL FRAMEWORK AND EMPIRICAL SPECIFICATION To explore the effects of different forms of insider/outsider ownership on employment adjustment and to test the above hypotheses, we start with a model of labor demand in the presence of adjustment costs. Adjustment costs in transition firms are expected to vary systematically by type of ownership. In a survey of labor markets in transition, Svejnar (1999) points out that most previous work specifies employment as a function of the real wage, output (measured by sales), and various ownership and other firm or industry specific dummies. In a dynamic cost-minimizing framework with adjustment costs, Nickell (1986) and Machin et al. (1993) derive Euler equations for dynamic labor demand that include lagged dependent variables to capture the costly process of partial adjustment as well as contemporaneous and lagged independent variables. This specification is essentially the general distributed-lag model specified by Basu et al. (1997, 2000) and Svejnar (1999), which nests various common, simpler specifications. We estimate this equation in differenced form using generalized method of moments (GMM).13 The resulting equation is n it = α0 + α1 n it−1 + α2 w it + α3 w it−1 + α4 Yit + α5 Dit + α6 Dit−1 + εit .

(1)

In (1), n it is the log of employment in firm i at time t, w it is the log of the real wage, Yit is the log of output proxied by sales and adjusted for inflation,  is the difference operator, and Dit is a set of dummy variables indicating stateowned, privatized, or other transitional forms of ownership observed in the Polish economy that may affect the demand for labor. This vector may contain other firmspecific characteristics available from our survey, such as unionization, change in management, percent of manual workers, and average firm size, that are relevant to 13 The lag structure has been determined empirically with a general-to-specific approach. Initially, we allowed for one lag on all variables and eliminated sequentially from the equations those with little explanatory power. See the empirical tables for details.

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labor demand.14 For estimation purposes, we include industry and time dummies to account for unobserved aggregate effects that may have influenced the economy or the firms in transition over time. In (1), εit is a composite error term in first differences including time-specific effects.15 The GMM estimator allows for the endogeneity of all regressors by using predetermined variables as efficient instruments. Using the same procedure for the wage equation to be specified below means that employment and wages are estimated implicitly as a system, in contrast to the procedure followed in much of the earlier transition literature.16 As Commander (1998), Grosfeld and Nivet (1999), and Svejnar (1999) explain, most firm-level transition studies estimate a wage equation in terms of sales per employee (labor productivity) and the usual vector of institutional and regional controls. Profits that enter formal bargaining models are either unavailable, as in our case, or unreliable. Thus, sales per employee are particularly important in the context of transition firms since labor productivity provides an upper bound to the rents workers might be able to extract and “is used to proxy for the firm’s ability to pay and hence the presence of rent sharing with workers” (Svejnar, 1999, p. 2830).17 We have no human capital variables such as skill and experience; hence, the lagged wage is used as a proxy for these variables, which change only slowly over time. The lagged wage is also included as a means of capturing partial adjustment in the dynamic estimation. The positive firm size–wage relationship has proved to be robust in Western firm-level studies, even when controlling for individual characteristics. Thus we include firm size, measured by employment, as an additional explanatory variable in our wage equation to test the firm size–wage effect for the first time in a transition context. In addition, we add regional alternative wages18 and unemployment rates as controls that influence employee bargaining power and allow for a direct test of how local supply and demand of labor influence wage outcomes, i.e., the well known wage-curve hypothesis (Blanchflower and Oswald, 1994, and Black and 14 Following Brown and Ashenfelder (1986), MaCurdy and Pencavel (1986), and Machin et al. (1993), we estimated another specification for equation (1) that included an alternative reservation wage (w ajt ) proxied by either average regional or industry regional wage to allow for bargaining over both wages and employment at the firm level. This yielded no significant results. See Basu et al. (2000) for an application to firms in early transition. 15 Note that the influence of time-invariant effects (η ) will be eliminated in first differences. i 16 Exceptions are K¨ ollo (1998), Commander and Dhar (1998), and Basu et al. (1999), who estimate two-stage least squares labor demand jointly with a wage equation. 17 It is worth pointing out the finding of Pinto, et al. (1993, p. 219): “. . . [in late 1992] . . . wage setting has come to resemble bargaining outcomes commonly seen in the West.” Thus, given the later stages of transition captured by our sample of firms this observation as argued above has come to motivate the approach to specifying our wage equation. 18 The alternative wage is defined as above but estimates are not reported for this variable because they provide little evidence of having a significant influence on wage outcomes. Grosfeld and Nivet (1999) examine the effects of the average sectoral wage in a similar context for a sample of Polish firms in early transition.

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FitzRoy, 2000). Since this hypothesis is usually stated in terms of wage levels, we specify two wage equations, one in levels and one in first differences. This allows us to compare, contrast, and complement our estimates and to test for the importance of partial adjustment to shocks in the dynamic specification. As before, the lag structure is determined empirically using the general-to-specific approach. The equations are w it = β0 + β1 w it−1 + β2 xit + β3 xit−1 + β4 n it   + β5 Un jt w ajt + β6 Dit + ξit w it = γ0 + γ1 w it−1 + γ2 xit + γ3 xit−1 + γ4 n it   + γ5 Un jt w ajt + γ6 Dit + ξit .

(2)

(3)

In these equations, the log of labor productivity is xit = Yit − n it and real sales are used as the output variable. The other variables and the error terms are defined as above.19 The new addition is the log of average unemployment rate for region j at time t (Un jt ). We have also estimated versions of the above equations with Un jt but it seems more plausible, both formally and intuitively, that the levels of regional unemployment determine wage outcomes in models of bargaining.20 Following the studies mentioned above, we expect productivity growth to be the main determinant of wage growth as it is in standard static models of rent sharing and bargaining. Empirically, productivity growth is a key component in wage negotiations. Grosfeld and Nivet (1999) use a similar specification for their wage equations, with the addition of the industry/sectoral average wage and regional unemployment rates. These authors emphasize the importance of controlling for firm fixed effects, but their Hausman test rejects fixed effects in favor of the randomeffects estimates that they report. Commander and Dhar (1998) use unemployment rates in their wage equation; however, none of the specifications they present yields a significant, nonzero, coefficient; i.e., there is no discernable effect of regional unemployment and local market conditions on wage growth. An obvious problem with the above analysis, which is not discussed by Grosfeld and Nivet (1999), is the endogeneity of regressors such as productivity. To address this issue, we report the first difference GMM estimates of our equations accounting explicitly for the endogeneity of the explanatory variables. Equations (1) and (3) follow the literature in assuming symmetric reactions to both demand/supply and productivity shocks. However, Commander and Dhar (1998), Svejnar (1999), and Grosfeld and Nivet (1999) suggest reasons to expect asymmetric responses; thus, in our empirical estimation, we distinguish between positive and negative shocks. Additionally, we expect differing reactions to 19

Note that Eq. (3) is not simply a differenced version of Eq. (2). Grossfeld and Nivet (1999) use a slightly different argument for early Polish transition and Black and FitzRoy (2000) provide an application to U.K. data. 20

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various shocks according to ownership categories. To implement this new approach, we distinguish between shocks by interacting the ownership dummies with the respective positive or negative sales and with productivity changes. Hence, another version of the employment equation can be written as n it = α0 + α1 n it−1 + α2 w it + α3 w it−1 + α4 YitSOE± + α5 YitNIF± + α6 YitPRII± + α7 YitPRIO± + α8 Dit + α9 Dit−1 + εit .

(4)

In (4), Dit excludes the ownership dummies but still contains the other controls; YitSOE± is an interactive term of the sales growth at time t for firm i if this growth rate is positive or negative, and if i is a SOE, 0 otherwise. Table 8 in the Appendix provides a detailed description of the variables and the acronyms used. Similarly for the wage growth equation, we have w it = γ0 + γ1 w it−1 + γ2 xitSOE± + γ3 xitNIF± + γ4 xitPRII± + γ5 xitPRIO± + γ6 n it± + γ7 Un jt + γ8 Dit + ξit .

(5)

This variety of empirical specifications allows us to test a number of different hypotheses discussed in Section 3. Our estimates may be seen as an extension and improvement of results provided in Basu et al. (1997) and Commander and Dhar (1998), as well as those in the studies mentioned earlier, in two important ways. First, our estimates are in first differences over a panel of 4 consecutive years and include a lagged dependent variable to capture the partial adjustment of endogenous variables at the firm level. Second, we employ the GMM estimators described below which correct for both the endogeneity, inherent in the properties of the explanatory variables, and the serial correlation properties of the residuals. For comparison, we have also estimated the wage equation in levels, following Basu et al. (1997, 2000). As a new variable, employment is included to test the wellknown firm size-wage effect, which turns out to be insignificant. However, in the differenced specification, which has also been used by Grosfeld and Nivet (1999), Commander and Dhar (1998), and Estrin and Svejnar (1998), our estimates reveal a significant effect of employment growth on wage growth. Our estimates assess the employment and wage decisions of Polish firms, paying close attention to the likely behavior of the insiders and outsiders in the process. We also expect pronounced differences with SOEs, as confirmed in earlier work. In the next section, we discuss the econometric issues that may arise in these equation specifications, describe the advantages of the GMM estimators used in this study, and report our findings. 5. EMPIRICAL METHODOLOGY AND RESULTS 5.1. Econometric Issues The panel structure of our sample allows us to study the dynamics of partial adjustment in the transition period by including a lagged dependent variable among the other regressors in the model. In such dynamic models with relatively large

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cross sections over a short time period, the fixed effects model yields inconsistent estimates due to the problem of incidental parameters. Thus, we specify an error components model with εit = λt + ηi + νit . In the presence of lagged dependent variables, well-known problems arise. Earlier work has used maximum likelihood estimators (MLE) and a simple instrumental variable (IV) approach (Bhargava and Sargan, 1983; Anderson and Hsiao, 1981) to address the issues of endogeneity and inconsistency. The relatively strong assumptions on the distributions of the individual effects, the initial conditions necessary to implement the MLE approach, and the lack of efficiency of the IV approach have encouraged the use of the Generalized Method of Moments (GMM) (Hansen, 1982) estimation in recent studies of dynamic panel regressions.21 We use the asymptotically efficient, one-step, GMM advocated by Arellano and Bover (1995) and more recently by Blundell and Bond (1998).22 This type of GMM estimator usually exploits a different number of instruments in each time period. Under weak assumptions, the additional orthogonality conditions that become available have not been previously used with IV estimators. Therefore, we should use transformations of the data that allow lagged endogenous or predetermined variables as instruments in the transformed equations, so that the transformed error term does not contain ηi and orthogonality among the errors is preserved. The original errors may be heteroskedastic but not autocorrelated; we treat all variables in our models as endogenous. An alternative approach to the first differences used here is the orthogonal deviations transformation (forward demeaning). When applied to our data, this approach yields similar results.23 To ensure consistency, we check for serial correlation in the errors. If εit are serially uncorrelated, εit = λt + νit may be moving average (MA) errors but should not be second-order serially correlated to assure the reliability of our results.24 However, Arellano and Bond (1991) note that, for T < 5, the test for second-order serial correlation (MA(2)) may not be well defined. Since T = 4 in our sample, only Sargan’s overidentification test is available from the two-step GMM estimation, which is heteroskedasticity-consistent. In our sample, there are 21 Baltagi (1995, Chap. 8) provides some background and a detailed discussion of GMM. Phillips and Moon (1999a, 1999b) contain an overview and treatment of the case when N and T are large, with the relevant asymptotic theory. 22 Earlier work by Arellano and Bond (1991) proposed the use of the IV approach in the GMM framework and employed the lagged levels of the series as instruments to the first differences. Ahn and Schmidt (1995) and Ahn and Schmidt (1997) discover additional moment conditions, which are nonlinear in the parameters, and show the efficiency gained in estimation. Using a similar approach, Hahn (1999) derives the sources of efficiency gains in the estimation method of Blundell and Bond (1998) and concludes that these can be substantial. In this connection, Blundell and Bond (1998) use simulations to suggest that, in typical sample sizes, the asymptotic standard errors for the one-step GMM estimators may be more reliable for inference. We report these in our empirical tables. 23 Arellano and Bover (1995) discuss other transformations with similar properties. 24 When ε are the first differences of serially uncorrelated errors, it is evident that E[ε ε it it it−1 ] may not be zero; however, for consistency, GMM estimators require that E[εit εit−2 ] = 0 (Arrelano and Bond (1991).

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no first-difference residuals that are two periods apart. Sargan’s test is a test of overidentifying restrictions, which is a chi-square under the null of no significance or instrument validity with the degrees of freedom (number of restrictions) given in parentheses in the empirical tables. For a simple first-order autoregressive (AR(1)) model and T = 4, Sargan’s test verifies the following three restrictions: E[(εi3 − εi2 )yi1 ] = E[(εi4 − εi3 )yi1 ] = E[(εi4 − εi3 )yi2 ] = 0, which in effect implies that the errors are not second-order serially correlated. The addition of regressors is done in a similar fashion; thus, we use MA(1) and Sargan jointly to determine the validity of our instruments and the correctness of our assumptions. 5.2. Employment Equations The first employment estimates in column 1 of Table 3 show the importance of allowing for partial adjustment, with a significant coefficient of about 0.7 on lagged TABLE 3 Estimated Employment Equation: Dependent Variable n it (164 Companies, Period 1994–1997, 328 Observations Available for Estimation) Independent variables n it−1 w it Yit Yit+ Yit−

(1)

(2)

0.681 (4.65) −0.021 (−0.27) 0.330 (2.21)

0.763 (6.47) −0.081 (−1.18) 0.058 (0.97) 0.355 (3.02)

Firm-specific variables % Trade union members Average firm size % Manual workers

0.001 (1.52) −0.025 (−2.58) 0.001 (1.22)

0.001 (1.53) −0.028 (−3.05) 0.001 (0.82)

Dominant ownership dummies NIF firm PRI–insider firm PRI–outsider firm

−0.030 (−1.84) 0.015 (0.84) −0.001 (−0.05)

−0.031 (−2.14) 0.023 (1.29) 0.005 (0.20)

Diagnostics MA(1) Sargan test Wald test for time and/or industry dummies

−2.136 12.02 (10) 49.83 (16)

−2.328 15.42 (11) 57.14 (16)

Note. (a) The equations are estimated by GMM in first differences. The t-statistic, reported in parentheses next to the point estimates, is corrected and robust to heteroskedasticity over industries and time. Constant and time dummies are always included but not reported; where appropriate, we add industry dummies and/or interactive industry/time dummies. The Wald test for the joint significance of those is reported in the last row of the table; the test is a chi-square under the null of no significance with degrees of freedom in parentheses. MA(1) is a test of first-order serial correlation, based on the standardized first-difference residual autocovariances asymptotically distributed as N (0, 1) under the null of no autocorrelation. Sargan’s test is a test of overindentifying restrictions, which is a chisquare under the null of no significance or instrument validity. (b) NIF, firms included in the National Investment Funds; PRI–insider firm, owned by insders; PRI–outsider firm, owned by outsiders.

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TABLE 4 Estimated Employment Equation: Dependent Variable n it (164 Companies, Period 1994–1997, 328 Observations Available for Estimation) Independent variables n it−1 w it SOE(+)

Yit

0.793 (7.77) −0.075 (−1.12) 0.074 (1.31)

SOE(−)

0.155 (2.13)

NIF(+)

−0.130 (−1.45)

NIF(−)

0.640 (4.11)

PRII(+)

0.156 (1.08)

PRII(−)

0.148 (1.07)

PRIO(+)

0.125 (1.14)

PRIO(−)

0.624 (2.88)

Yit Yit Yit Yit Yit Yit Yit

Firm-specific variables % Trade union members Average firm size % Manual workers

0.001 (1.64) −0.030 (−3.38) 0.001 (1.00)

Diagnostics MA(1) Sargan test Wald test for time and/or industry dummies

−2.178 15.59 (11) 54.33 (16)

Note. See Table 3 note.

employment growth.25 The change in contemporaneous sales has a significant effect on employment changes as expected and found in earlier studies. From column 2 of Table 3, the significant asymmetric response only to negative sales growth only reveals that sales increases have had little or no effect on employment growth. In other words, employment growth responds strongly only to decreasing sales, other things being equal. This interesting result suggests that labor hoarding was initially widespread so that firms can raise output, or respond to positive demand shocks, with little additional labor. Estrin and Svejnar (1998) obtain a similar finding using a different specification in an earlier sample of Polish firms. Furthermore, declining sales or adverse demand shocks do prompt significant downsizing by the firms in our sample. We find additional, significant asymmetric responses to sales growth in Table 4 in which we interacted ownership dummies with positive and negative growth rates of sales to allow for asymmetric demand shocks. None of the estimates suggests a significant positive employment response to positive sales growth, confirming again the labor hoarding hypothesis. 25 The coefficients of lagged employment growth may easily be shown to be statistically different from one.

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From Table 4, SOEs exhibit the smallest (0.16) significant response to negative shocks. NIFs and PRIOs do appear to downsize significantly in response to negative sales shocks. The result for NIFs is indicated in Table 3 where ownership dummies seem to affect significantly employment change for only this category of firms. However, earlier studies suggest that comprehensive restructuring does not occur in mass privatization programs due to ownership uncertainty (Aghion and Carlin, 1997). Our results point in the opposite direction in that employment adjustment takes place as a result of a significant negative shock to sales in NIFs and the response is slightly stronger than it is in PRIOs.26 Not surprisingly, PRIIs do not respond by adjusting employment to sales growth. There is no asymmetric response to sales growth since neither of the coefficients is significantly different from zero. This is consistent with the hypothesis of insiderdominated firms translating positive demand shocks into higher wages rather than higher employment.27 However, given sales, we do not find evidence for the expected negative relationship between wage growth and employment in our sample of firms as the coefficients on w it in Tables 3 and 4 indicate.28 Finally, other things being equal, we confirm that larger firms tend to grow slower, since the coefficient for average firm size is negative and significant in Tables 3 and 4. However, we are not able to find a significant union effect on employment adjustment; in unreported results, we fail to establish the presence of a firm-size union effect. The time-dependent, firm-specific variable, measured by the percent of manual workers, is not significantly different from zero, suggesting that employment adjustment is not related to job skills. Diagnostics show that neither the robust Sargan nor MA(1) test provides evidence to suggest that the assumption of serially uncorrelated errors (second-order) is unrealistic. According to these tests, the choice of the instruments is appropriate. These results also hold for the wage-equation estimates to which we turn next. 5.3. Wage Equations The wage growth specification in Table 5 shows significant effects of both productivity growth and employment growth, as revealed by the coefficients for xit and n it , respectively. The lagged dependent variable is significantly different from zero but smaller in size than the above effects, suggesting that wage changes 26 We thank an anonymous referee for suggesting this line of argument to us. Although not reported NIFE(−) PRIO(−) and Yit here, we conducted a Wald test for the joint significance of the coefficients of Yit and drawing attention to their relative size is warranted. 27 See Blanchard and Aghion (1996) for a theoretical perspective. In the early transition period from 1990 to 1991, Estrin and Svejnar (1998) find similar results for Polish cooperatives, a group not covered in our study. 28 Earlier studies have found negative employment-to-wage-growth elasticities that increased over time and became more pronounced as the transition proceeded. However, in unreported estimates of employment levels available upon request, in which we include a lagged dependent variable, the wage (level) was negative and significant.

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TABLE 5 Estimated Wage Equation: Dependent Variable w it (164 Companies, Period 1994–1997, 328 Observations Available for Estimation) Independent variables w it−1 xit xit+ xit− n it n + it n − it

(1)

(2)

(3)

0.380 (2.88) 0.596 (4.06)

0.365 (2.77) 0.619 (4.12)

0.353 (2.65) 0.672 (3.25) 0.501 (2.21)

0.740 (3.31) 1.214 (2.07) 0.560 (2.64)

1.176 (1.96) 0.594 (2.60)

Firm-specific variables Un jt (Regional) % Trade union members % Manual workers % Female workers

−0.095 (−1.92) 0.002 (1.68) −0.004 (−2.23) −0.003 (−1.60)

−0.092 (−1.90) 0.002 (1.78) −0.005 (−2.33) −0.002 (−1.56)

−0.091 (−1.85) 0.002 (1.81) −0.005 (−2.32) −0.003 (−1.57)

Dominant ownership dummies NIF firm Privatized-insiders Privatized-outsiders

−0.011 (−0.22) 0.080 (1.29) 0.076 (1.18)

−0.011 (−0.23) 0.086 (1.42) 0.065 (1.03)

−0.006 (−0.13) 0.093 (1.53) 0.068 (1.07)

Diagnostics MA(1) Sargan test Wald test for time and/or industry dummies

−3.334 14.54 (16) 84.05 (9)

−3.289 13.93 (16) 84.88 (9)

−3.238 13.18 (16) 71.92 (9)

Note. See Table 3 note. Un jt , log unemployment rate in region j at time t, 49 voivodships from 1994 to 1997.

are not highly autocorrelated but rather driven primarily by productivity growth as expected, with the additional new finding of a large effect of employment expansion. Interestingly, positive employment changes have a much larger effect on wage growth than employment declines (see rows six and seven of columns 2 and 3 of Table 5).29 This result, together with the insignificant coefficient of wage levels with respect to firm employment discussed below, suggests that this employment growth-wage effect may be due mainly to the wage increases needed to attract new employees from existing jobs or from beyond the local labor market to growing firms. Thus, information asymmetries and costs of labor mobility generate an upward-sloping labor supply curve or a temporary dynamic monopsony in contrast to the traditional competitive labor market model. However, the smaller effect of employment declines on wage growth is consistent with efficiency wage theories and the traditional resistance of workers to wage moderation even when employment is falling. 29 Note that, by construction, the variable n − is negative and, hence, a positive elasticity coefficient it indicates a reduction in the growth rate of wages.

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CHRISTEV AND FITZROY TABLE 6 Estimated Wage Equation: Dependent Variable w it (164 companies, Period 1994–1997, 328 Observations Available for Estimation) Independent variables w it−1

0.380 (3.30)

SOE(+)

0.300 (1.71)

SOE(−)

0.409 (1.78)

NIF(+)

0.206 (0.62)

NIF(−)

1.367 (1.83)

PRII(+)

1.823 (5.44)

PRII(−)

1.094 (1.49)

PRIO(+)

0.608 (1.80)

PRIO(−)

0.432 (1.22) 1.300 (2.13) 0.471 (2.16)

xit xit xit xit xit xit xit xit

n + it n − it Firm-specific variables Un jt (Regional) % Trade union members % Manual workers % Female workers

−0.071 (−1.45) 0.002 (2.01) −0.006 (−2.84) −0.003 (−1.58)

Diagnostics MA(1) Sargan test Wald test for time and/or industry dummies

−3.543 14.75 (16) 79.89 (9)

Note. See Table 3 note. Un jt , log unemployment rate in region j at time t, 49 voivodships over 1994–1997.

To our knowledge, this asymmetric response has not been found in empirical studies of transition firms. The specification of the wage equation that we prefer is column 3; it indicates distinct effects of positive and negative productivity shocks. Wage growth is more responsive to positive than to negative productivity shocks, other things being equal.30 Finally, in contrast to earlier work summarized in Svejnar (1999), we provide evidence of a wage-curve effect, because the coefficients of regional unemployment are negative and significant in all three equations in Table 5. Turning to the interactions of ownership dummies with positive and negative productivity shocks (Table 6), some striking asymmetries emerge for all firms. Interestingly, SOEs respond to negative productivity shocks by reducing wage growth; however, we cannot reject the hypothesis that these positive and negative productivity effects are identical for this type of firm. On the other hand, negative 30

A Wald test rejects the hypothesis that the two coefficients are identical.

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productivity shocks have a significant and large moderating effect on wage growth in NIFs. NIFs are the only category of firms that show no significant effect of productivity increases on wage gains; since productivity rises were not translated into wage increases, substantial restructuring may have occurred contrary to the claims in the earlier literature. Together with our employment results, this finding implies that NIFs responded to transition shocks by adjusting both employment and wages. While negative shocks do not have a significant effect on wage growth, positive productivity shocks have a powerful and significant influence on PRIIs’ wage gains. Taking these coefficients as a measure of insider power as suggested by Grosfeld and Nivet (1999), insider-controlled firms exhibit the highest positive productivity elasticity with respect to wage growth. This elasticity is three times as high as that for PRIOs. Finally, positive employment growth has a significant coefficient twice as large as that for negative employment change. Again, this implies that the large employment growth-wage effect is due to growing firms raising wages to attract workers in temporary dynamic monopsony situations. In the tables, two firm-specific variables have a significant effect on wage growth. Not surprisingly, trade unions have a small positive influence and manual workers (defined as unskilled labor and comprising 70% of the work force in the sample on average) have a negative impact on wage growth in Tables 5 and 6. However, the results in Table 6 do not confirm the wage-curve hypothesis as the coefficients on regional unemployment are insignificant. This rather unexpected finding points to the relative importance of own productivity shocks for firms and the somewhat minor impact of local labor market conditions in the transition thus far. To corroborate our findings, wage estimates in levels in Table 7 show that the lagged wage dominates the other explanatory variables, reduces the productivity influence, and shows no size-wage effect, in contrast to our more informative wage growth specifications in Tables 5 and 6. In line with Basu et al. (1997) and the other literature surveyed in Svejnar (1999), we find the wage elasticity with respect to labor productivity to be 0.5 for the general distributed lag model (column 2 of Table 7).31 These estimates do not support a wage-curve effect; i.e., we do not find a significant influence of regional unemployment rate on wages. Except for insider-controlled firms that raise the level of wages significantly, other ownership categories seem to have no influence. These results confirm the importance of our differenced specifications that do reveal additional effects missed by earlier studies. 31 We have included two additional firm-specific, time-invariant variables, namely change in management and indicated profitability, which our differenced specification excludes by construction. For ease of comparison, we do not report these results. However, such a specification yields productivity estimates of 0.3 to 0.4, which are similar to those in earlier studies surveyed in Svejnar (1999), and provides evidence of a significant effect of the change in management on wage formation. This result is consistent with the findings of Pinto et al. (1993) for firms in the early transition period in Poland.

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CHRISTEV AND FITZROY TABLE 7 Estimated Wage Equation: Dependent Variable w it (GMM Estimates: Period 1994–1997, 492 Observations Available for Estimation) (1) IV estimates

(2) System GMM

0.838 (4.69) 0.556 (3.83) 0.067 (0.58)

0.834 (14.54) 0.544 (5.18) 0.070 (0.91) 0.014 (0.12)

Firm-specific variables: Un jt (regional) % Trade union members % Manual workers % Female workers

−0.183 (−0.99) 0.003 (0.74) 0.009 (1.41) −0.013 (−2.27)

−0.064 (−1.16) 0.002 (1.56) 0.005 (1.74) −0.001 (−0.41)

Dominant ownership dummies NIF firm Privatized-insiders Privatized-outsiders

−0.126 (−0.68) 0.540 (2.93) −0.047 (−0.22)

−0.032 (−0.36) 0.145 (1.75) −0.048 (−0.49)

Independent variables all firms (164 firms) w it−1 xit xit−1 n it

Diagnostics MA(1) Sargan test Wald test for time and/or industry dummies

−4.316 22.59 (23) 119.30 (9)

Note. See Table 3 note. System GMM estimates are obtained by stacking (T − 2) equations in first differences and in levels corresponding to periods 3, . . . , T . We then use lagged differences of the variables as instruments in levels ((w, Y , x, Un) dated t − 1, etc.) in addition to the instruments specified for the difference equations (see Blundell and Bond, 1998, p. 126).

6. CONCLUSION Our results provide evidence of initial widespread labor hoarding in Polish firms because employment growth responds strongly only to decreasing sales. Falling demand has prompted substantial downsizing in all but the insider-controlled firms. Our evidence suggests that, although PRIIs do not appear to adjust employment much to sales growth, NIFs reduce employment substantially in response to negative output shocks. This indicates that, despite initial fears, these firms have restructured substantially in anticipation of final ownership concentration. We also provide evidence that insider-controlled firms appropriate rents by converting any positive productivity shocks into large wage gains, unlike other firms in the sample. Finally, we do not find a significant firm size-wage effect, a common empirical regularity in Western firm-level data but hitherto overlooked in transition enterprises. However, our wage growth equation reveals an employment-growth effect twice as large for increasing-employment firms than for labor-shedding ones. This suggests that we have identified labor supply since growing firms raise wages to attract workers in temporary dynamic monopsony situations.

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APPENDIX: DATA The survey was completed early in 1998 with the support of the EU-ACE Program and partner-institutions in Poland. In the process, detailed questionnaires were sent to firm managers and interviews were conducted with a representative of the accounting office in most of those enterprises. The questionnaire is divided into four parts. The first part consists of questions concerning location and activities of the enterprise, description of the ownership type and its changes, and quantitative managerial data when available. The second part is about the employment structure and policy of the company. The third part deals with the influence of state subsidies and regulations on the enterprises regarding employment and investment decisions. The last part includes qualitative information on the voluntary and compulsory social benefits provided by the company. Hence, the survey data contain information on employment, wages, and other firm characteristics for the period; we use these data to construct our panel. Trade unions were active in 88.7% of the enterprises surveyed, with two or three different unions operating in some enterprises. Positive profitability over the sample period was indicated by 77% of our firms. Competition was viewed as strong in 58.4%, while 27% of the enterprises produced for the domestic market only. Changes in management were observed in 40.4%. As a sign of hardening budget constraints, 84.3% received no subsidies from the state from 1994 to 1997. While only 13.5% were allowed to accumulate tax arrears, no subsidies were provided by the state for wages or social benefits paid out by the enterprise. Given the diversity of the enterprises, the managers’ perceptions of the government’s influence on employment and wages were captured in a sequence of qualitative questions that required the managers to rank the influence from nonexistent to strong. With regard to wages, 35.4% of the managers considered the influence of the Tripartie Commission (comprised of unions, government, and employers) to be strong, 46.1% conceded the pressure exerted by trade unions and employees to be strong, and 73.6% reported that cost-effectiveness had a strong effect on wage setting. Regarding employment, 60.7% of the managers perceived the government’s influence as nonexistent and 33.7% of them reported that its strength had decreased since 1994. The industries used in the study include mining and quarrying (excluding gas and coal); manufacturing of food products; beverages; textiles; wearing apparel and furriery; leather and leather products; wood and wood products; pulp and paper; chemical and chemical products; rubber and plastic products; other nonmetal mineral products; basic metals; not classified machinery and equipment; not classified electrical machinery and apparatus; radio, TV, and communication equipment; medical, optical, and precision equipment; motor vehicles and trailers; furniture and other manufacturing; electricity, gas, and hot water supply; construction; wholesale and retail trade; and other land transport (excluding railways). Table 8 provides a description of the variables and acronyms used throughout.

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CHRISTEV AND FITZROY TABLE 8 Identification of Variables Used in the Text Variables

Employment n it n + it n − it Wages w it

Sales Yit Yit+ Yit− SOE(±)

Yit

NIFE(±)

Yit

PRII(±)

Yit

PRIO(±)

Yit Labor productivity xit xit+ xit−

SOE(±)

xit

NIFE(±)

xit

PRII(±)

xit

Description

Growth rate of log employment of firm i at time t; average growth rate (−0.048) Positive growth rate of log employment of firm i at time t, 0 otherwise Negative growth rate of log employment of firm i at time t, 0 otherwise Growth rate of log real product wages (the wage deflated by PPI1993=100 : 1994 = 125.3, 1995 = 157.1, 1996 = 176.6, 1997 = 198.2); average growth rate (0.472) Growth rate of log real sales/turnover of enterprise i at time t; average growth rate (0.016) Positive growth rate of log sales of firm i at time t, 0 otherwise Negative growth rate of log sales of firm i at time t, 0 otherwise Interactive variable, Yit± multiplied by SOE dummy

Interactive variable, Yit± multiplied by NIFE dummy Interactive variable, Yit± multiplied by PRII dummy

Interactive variable, Yit± multiplied by PRIO dummy Growth rate of log labor productivity, defined as xit = Yit − n it ; average growth rate (0.064) Positive growth rate of log productivity of firm i at time t, 0 otherwise Negative growth rate of log productivity of firm i at time t, 0 otherwise Interactive variable, xit± multiplied by SOE dummy

Interactive variable, xit± multiplied by NIFE dummy Interactive variable, xit± multiplied by PRII dummy

xit Interactive variable, xit± multiplied by PRIO dummy Time variant and firm specific Log unemployment rate in region j at time t, 49 voivodships over Un jt 1994–1997 % Trade union members Average percentage of trade union membership 1994–1997; average value 47% (maximum 85%) % Female workers Time-dependent percentage rate: highest mean value over 1994–1997, 41.5% of labor force in 1995 (maximum 95%) % Manual workers Time-dependent percentage rate: highest mean value over 1994–1997, 75% beginning of period (maximum 95%) Dummy variables SOE Dummy variable state-owned enterprises = 1, 0 otherwise NIF Dummy variable national investment fund firm = 1, 0 otherwise PRII Dummy variable privatized firms-insiders = 1, 0 otherwise PRIO Dummy variable privatized firms-outsiders = 1, 0 otherwise Indicated profitability Dummy variable, 1 if firm indicated profits over 1994–1997, 0 otherwise Change in management Dummy variable, 1 if firm management has changed over 1994–1997, 0 otherwise PRIO(±)

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