Jose Alberto Molina
email: jamolina@unizar.es

Inmaculada Garcia

Victor Manuel

Alberto Molina

Manuel Montuenga

José Alberto Molina

Intra-family distribution of paid-work time

Keywords: Intra-family, Paid-work time, Collective model Intra-family, Paid-work time, Collective model

INTRODUCTION

In recent times, we have witnessed an increasing amount of research effort devoted to pursuing an adequate theoretical framework for modelling the process of intra-family decision-making. A review of the literature appears to confirm that the traditional or unitary approach, which assumes that a household, even if it consists of different individuals, acts as a single decision-making unit, is giving way to an alternative view, which considers that a household can be seen as a micro-society consisting of several individuals with their own rational preferences. 1 This evolution is due to the fact that the unitary approach suffers from a number of both methodological and empirical weaknesses. Thus, from the methodological perspective, the idea that subjective preferences are inseparable from individual behaviour leads to an alternative approach, which explicitly takes into account the notion that a household is a group of individuals, with different preferences, and among whom an intra-family decisionmaking process takes place. As regards empirical weaknesses, the most relevant appears in the context of the welfare analysis, since the unitary model leaves no room to determine the intra-family distribution of consumption or time and, consequently, of welfare. 2 As a result, knowledge of the intra-family decision process may be important for the design of policy programs whose aim is to contribute to the development of family members and, in this way, of the household as a whole.

In response to these weaknesses, two fruitful approaches have emerged in the literature which explicitly take into account several decision-makers in a household by 1 Samuelson (1956) and Becker (1974a, 1974b) represent the early attempts to account for the fact that households may consist of different individuals with their own preferences. However, in both cases the authors finally accepted the traditional approach: in the first case, through an aggregate utility function which is achieved by consensus among the individuals; in the second, by assuming the utility function of a benevolent head of the family, who takes into account the preferences of all household members.

2 Other empirical restrictions, strongly rejected in the literature, include the income pooling hypothesis and the symmetry of the Slutsky matrix (see, among others, Kooreman and Kapteyn, 1987; Schultz, 1990; Thomas, 1990; Fortin and Lacroix, 1997; Tiefenthaler, 1999). With respect to the former, this implies that individual non-labour incomes of the household members are pooled in a single household non-labour income which, in turn, implies that the source of this exogenous income plays no role in the household's distribution. As regards the latter, this requires that marginal compensated price changes of two individuals in a household have the F o r P e e r R e v i e w 3 using game theory elements. The first of these models household behaviour in a noncooperative framework, in such a way that the Nash equilibrium implies that family members are assumed to maximise their utility, taking the other individuals' behaviour as given (see, for example, Weiss and Willis, 1985; Kooreman and Kapteyn, 1990; Konrad and Lommerud, 1995; Browning, 2000). The second incorporates elements of co-operative game theory in a household model, specifically that of axiomatic bargaining theory, in such a way that household members reach the Nash or the Kalai-Smorodinsky solutions after trying to come to an agreement on how to divide the gains of co-operation (see Manser and Brown, 1980; McElroy and Horney, 1981, McElroy, 1990). Despite the clear relevance of these two game theory approaches, one important drawback they share is that if the empirical implications of such approaches are rejected, then it is impossible to determine whether the choice of a particular bargaining concept itself, or the bargaining approach in general, is the cause of such rejection.

Against this background, Chiappori (1988, 1992, 1997) adopts, within the cooperative view, an alternative, and gradually more accepted approach, namely the collective model, which is based on the assumption that intra-family decisions are Pareto-efficient, thus making it possible to derive some testable implications and to identify an important part of the intra-family decision-making process. Several reasons justify the adequacy of this approach. First, it is reasonable to assume that the result of a repeated game under perfect information on each other's preferences within a household is Pareto-efficient; secondly, this is the most natural generalisation of the utility maximisation in the unitary model with several household members; and, thirdly, most of the commonly applied bargaining rules usually assume this view. The collective approach considers that preferences depend on wages, prices and individual non-labour incomes, in such a way that the distribution of the bargaining power within a household may depend on the level of each of these variables. Moreover, this alternative approach overcomes the two fundamental empirical restrictions of the unitary model, that is to say, the income pooling hypothesis and the symmetry of the Slutsky matrix, and can be also seen as a more general model which encompasses the unitary model. Bearing in mind that the collective model allows us to test for such a setting through testable implications derived from it and then to estimate the structural model from observed behaviour, the aim of this paper is, first, to test the adequacy of the collective approach in five representative EU countries (France, Germany, Italy, Spain and the UK) and, in the positive case, to analyse the intra-family distribution of paid-work time by estimating labour supply functions.

After transforming the two-equation paid-work time model into its empirical counterpart, we simultaneously estimate this specification by using national panel data drawn from the European Community Household Panel-ECHP (1994-2001). The structure of the panel, which includes relevant information on the factors that affect the alternative uses of household time, specifically those related to working outside the home in a paid job, allows us to control the unobservable heterogeneity problem, as well as to eliminate the bias derived from aggregation.

In order to fulfil the identification and testability requirements of the collective approach, we follow Chiappori et al. (2002) in considering the existence of distribution factors. These are variables that may have an influence on family behaviour through their effect on the intrahousehold decision process, but which do not affect either individual preferences or the household budget constraint. Some examples used in the literature are the sex ratio, divorce laws, differences in incomes, in ages or in education levels between both spouses. As discussed below, in our particular case, we use the share of the wife's nonlabour income over the household non-labour income.

Our paper sets out to make three main contributions. First, we present evidence about the collective model of household labour supplies for a set of countries using the same database, thereby allowing us to make a comparative analysis of the intra-family distribution of paid-work time in European economies. The second contribution is the use of techniques related to panel data with the aim of controlling for unobserved heterogeneity.

At the same time, endogeneity arising from simultaneity is also considered in the estimation process. Finally, the value of the estimations based on this collective framework lies in obtaining evidence about the family characteristics that influence decisions on the hours The rest of the paper is organised as follows. Section II presents the theoretical framework and the empirical specification. The data and the estimation procedure are described in Section III. The results and policy implications are discussed in Section IV.

Finally, Section V closes the paper with a summary of the most relevant conclusions.

II. THEORETICAL FRAMEWORK

II.1. The collective model

The traditional approach of a unitary model, which assumes that a family acts as a single decision-making unit, has given way in the literature to an alternative view which considers that a household can be seen as a micro-society consisting of several individuals with their own rational preferences. Within this framework, Chiappori and his co-authors (Chiappori, 1988, 1992; Browning et al., 1994; Browning and Chiappori, 1998; Chiappori et al., 2002) propose an approach that has gradually gained more acceptance in both theoretical and empirical applied work. 3 The so-called collective model, which is based on the assumption that intra-household decisions are Pareto-efficient, considers that the household consists of two working-age persons, A = husband and B = wife, whose rational preferences can be represented by individual utility functions that, in general, are assumed to be altruistic, in such a way that each of them is defined on their own vectors of goods and time, as well as on the other member's vectors:

( ) i i A B A B 0 0 u u q ,q ,q ,q , = z (1)
where u i , i A,B = , are strongly quasi-concave, increasing and twice continuously differentiable functions. The arguments are the consumption of each of the spouses, A q and 3 See Vermeulen (2002) for an excellent recent survey on the collective household approach. q , whose prices are unity, as well as their leisure times A q 0 and B q 0 ,4 with z being a vector of preference variables including family characteristics. Furthermore, the household budget restriction is:

( )

A B A A B B A B 0 0 q q q q y T + + + + + ( 2
)
where i denotes the individual wages, y is the family non-labour income and, finally, T is the time endowment.

According to the collective approach, the household demand functions can be derived from an intra-family decision process whose only requirement is that it must lead to Paretoefficient allocations, with such a process being formally implemented in the following maximisation problem:

( )

A B A B 0 0 A A B A B 0 0 q ,q
,q ,q Max u q ,q ,q ,q ,z s. to ( )

B A B A B B 0 0 u q ,q ,q ,q , u z (3) ( ) A B A A B B A B 0 0 q q q q y T + + + + +
where B u is some required utility level for individual B. From this initial problem, B u can be modified in order to obtain all the Pareto-efficient allocations, with these forming the boundary of the utility possibility set.

Given the initial assumption that the individual utility functions are strictly quasiconcave, and that the budget restriction defines a convex set, the utility possibilities set will be strictly convex. Consequently, all the Pareto-efficient allocations can be characterised as points of a linear social welfare function with positive weights for both household members in the joint welfare. Thus, the above problem can be expressed in the following terms:

F o r P e e r R e v i e w 7 ( ) ( ) ( ) ( ) A B A B 0 0 A B A A B A B A B B A B A B 0 0 0 0 q ,q ,q ,q

Max

, ,y, , u q ,q ,q ,q , 1 , ,y, , u q ,q ,q ,q , µ µ + s z z s z z (4) s. to ( )

A B A A B B A B 0 0 q q q q y T + + + + +
where s is a vector of distribution factors. In this optimisation problem, the weights

( ) A B
, ,y, , µ s z and ( ) ,y,,µ s z are the (normalised) Lagrangian multipliers of problem (3), with these being interpreted as indicators of the bargaining power of the household members in the intra-family distribution process. 5 They are assumed to be continuously differentiable and homogeneous of degree zero in y and w. As can be appreciated from these expressions, the collective setting implies that the bargaining power depends on the individual wages, ( )

A B 1 ,
A B
, , the household non-labour income, y, the set of family characteristics, z, and the distribution factor, s. These latter distribution factors denote variables that have an influence on family behaviour through their effect on the intrahousehold decision process (i. e. on the bargaining power function µ), but that do not affect either an individual's preferences or the household budget constraint (Browning et al. 1994; Browning and Chiappori, 1998).

The main limitation of this general setting derives from the fact that the structural model is not uniquely identified with altruistic preferences, in such a way that additional assumptions are required in order to identify the structural model. Note, however, that the integrability requirements do hold under both egoistic and caring à la Becker preferences (Chiappori, 1988, 1992). As a consequence, in what follows, we will consider the first of these in order to simplify the notation, although the results are easily extended to the case of the most general Beckerian utility functions.

Thus, assuming that family members only care about their own decision variables, ( )

i i i i 0
u u q ,q , = z , the second fundamental theorem of welfare shows that the efficient solution derived from this optimization problem is equivalent to a two-stage decision process (Chiappori, 1992). In the first stage, both family members share the household nonlabour income according to an exogenous and unobservable sharing rule which characterises the bargaining process within the household:

( )

A A B , , y, = s, z ( ) B A B A B y y , , y, = + s, z
where i represents the fraction of the household non-labour income that each member receives. The share depends on the bargaining power of the household members which, in turn, depends on wages, total non-labour income, sociodemographic variables and distribution factors. It is assumed to be twice continuously differentiable. Once the total non-labour income has been allocated between the individuals, in the second stage each family member maximises his/her own utility function, subject to his/her own budget restriction:

( ) i i 0 i i i i 0 q ,q i i i i i 0 Max u = u q ,q , s.to q + q + T ዊp z
Assuming an interior solution and provided that individual preferences are assumed to be weakly separable in 0 ( , )

i i q q , the individual labour supply functions of each spouse can be written as:

( ) ( ) ( ) A A B A A A A B h , , y, , H , , , y, , = s z s z ( ) ( ) ( ) B A B B A A B h , , y, , H , y , , y, , = s z s z
power. This means that the income pooling hypothesis no longer needs to be true. where i H are the Marshallian labour supply functions corresponding to the second stage of the problem and, therefore, are conditional on the sharing rule.6

In this framework, Chiappori et al. (2002) show that it is possible to derive testable restrictions on observed individuals' labour supply behaviour, and to recover the sharing rule function up to an additive constant. Despite that, even without distribution factors, Chiappori (1988, 1992) demonstrated that, under egoistic or caring preferences, the model is identified and the testable requirements are satisfied. The inclusion of distribution factors provides a simpler and more robust method of identification and the derivation of testable restrictions relies on first and second-order partial derivatives of the individuals' labour supplies. Several distribution factors have been used in applied work, with differences in income between spouses having been the benchmark (Browning et al., 1994). More

recently, Chiappori et al. (2002) used the state sex ratio and the divorce legislation; Clark et al. (2002), in addition to the regional sex ratio, employed other factors, such as the parents' occupational level or variables measuring political and sociological involvement; Crespo (2005) used, among others, the differences in education level between spouses. Given that only one distribution factor is needed for the testability and integrability requirements (Chiappori et al., 2002), in this work we consider that the intrahousehold decision process may be affected by the amount of non-labour income that each spouse pools to the household income. Specifically, the distribution factor we use is the fraction of the wife's non-labour income over the household non-labour income, with this variable influencing both members' labour supply behaviour through the sharing rule, but without affecting either individual preferences or the household budget constraint.

II.2. Parametric specification

In order to derive the testable restrictions, assumptions about the functional form of either the utility function or the labour supply functions must be imposed. Different empirical specifications have been used in applied work (see Fortin and Lacroix, 1997; Chiappori et al., 2002; Fernández-Val, 2003; Bloemen, 2004). From among those used in this literature, we have chosen both semilog and quadratic parametric representations that allow us to test the restrictions imposed by the collective framework on the observed labour supplies. These particular functional forms for the unrestricted labour supply system satisfy two widely followed criteria, namely flexibility in response to wage changes, and the possibility of recovering the sharing rule.

The semilog specification is fully described in Chiappori et al. (2002), with its main advantage being that the log form for the wages is more realistic than the habitually applied "pure" linear form:

A 0 1 2 3 4 5 A B A B 6 h log log log log y = + + + + + + s z 0 1 2 3 4 5 6 B A B A B h log log log log y = + + + + + + s z
The application of the general conditions derived in Chiappori et al. (2002) to these two unrestricted parametric models allows us to determine that there is only one parametric restriction:

( ) =
and k is a parameter that cannot be identified without additional assumptions (see Chiappori et al., 2002 The application of the same general conditions as above allows us to determine that, once again, there is only one parametric restriction: 3 3 7 7 = . As was earlier the case, if this restriction is empirically satisfied, we can obtain the following expression for the sharing rule:

3 1 2 3 3 3 3 4 5 3 3 7 3 A B A2 B 2 A B A B 6
( , , y, , )

= + + + + + + + s z s where 6 3 3 6 ( ) = 1 ( y ) k =
and k is a parameter that cannot be identified without additional assumptions.

III. EMPIRICAL MODEL AND DATA

III.1. Empirical model

In order to develop the empirical specification, a number of considerations should be made.

First, given that the ECHP data base does not include information with respect to the time spent on housework, we have had to include this in our leisure time, an activity not considered in our analysis. Secondly, the panel data structure allows us to apply techniques that help to control for unobservable heterogeneity. In this respect, there are three potential sources of unobservable heterogeneity: the sharing rule and the two individual preferences. The most satisfactory treatment would be to allow for each of them and then to develop a full stochastic model that would also allow us to take account of the possible endogeneity of the sample selection of married couples in full-time employment (Browning et al., 1994).

However, following these authors and Blundell et al. (2002) we adopt a much more conventional approach of simply adding error terms to each labour supply equation, allowing for individual effects to control for unobservable heterogeneity but ignoring the possible sample selection bias. 8

Thirdly, the distribution factor, s, considered in this analysis is given by a variable that reflects the share over the household non-labour income that corresponds to the wife, while vector z gathers a series of a priori exogenous variables that influence individuals' decisions. 9 The rationale for the distribution factor is that a higher share allocated by the wife indicates a higher relative power in the decision-making process. This will imply, conditioned on the wage rate, a lower number of hours of work for the wife as a result of a standard income effect, under the assumption that leisure is a normal good. The influence of this factor on the husband's labour supply function will have the opposite sign, according to the same argument.

In line with the above, the empirical labour supply functions we estimate take the following semilog or quadratic forms:

8 This supposes we are not considering the possibility of nonparticipation in the labour market and then selectivity bias may arise. Rather, we focus on the case of the interior solutions. The main reasons why we have opted to ignore this bias in our analysis are the difficulties in both considering corner solutions in the collective model (see Blundell et al., 2002; Donni, 2003; and Bloemen, 2004) and, similarly, dealing with selectivity bias in the case of panel data estimation (see Wooldridge, 2002, Kyriazidou, 1997). The consideration of all these possibilities is beyond the scope of this paper and is left for future research. 9

As stated earlier, different distribution factors have been used in applied work. Although we have tried a number of these, we finally chose this one given the inadequacy or non-significance of other possibilities. Thus, divorce laws are common within a country, the low regional disaggregation in the database used makes the construction of appropriate sex ratios more difficult, while differences in ages, in educational levels or in experiences between spouse are not applicable with the fixed effects estimation, which was ultimately selected, since they are time-invariant. Finally, this is a potentially continuous function and, thus, does not give rise to any incompatibility with the required assumption of continuously differentiability of the sharing rule, which usually fails elsewhere (Chiappori et al., 2002; Clark et al., 2002; Crespo, 2005).

Page

A B , =
is the vector of error terms that include individuals' unobservables, possibly correlated within the household, but not across households

At this point, it should be mentioned that there are some econometric concerns that must be properly dealt with. The ECHP provides annual earnings and weekly hours worked, so that hourly wage rates are computed by dividing annual earnings coming from paid-work by the number of hours worked in a week times the number of weeks in a year. This computation may generate some measurement error in the wage variable. This fact, and the simultaneity between hours worked and wages, leads us to regard the latter as endogenous, and then to follow an instrumental variables estimation. By contrast, family non-labour income and the children variable will be considered as exogenous. In order to deal with the endogeneity of wages, we have carried out a two-stage estimation. In the first, we estimate two wage equations in which each member's observed wage is regressed in a set of exogenous explanatory variables (experience, educational attainment, occupation, firm size, the type of sector -public or private-and the type of contract -fixed-term or permanent) and, subsequently, we construct an "exogenous" measure of individual wages which are included in the estimation of the system of equation of hours worked. (1994-2001). 10 The EHCP contains data on individuals in the fifteen pre-enlargement EU countries. It provides abundant information about both the personal and labour characteristics of individuals. This information is homogenous across countries, as the questionnaire is the same and the elaboration process of the survey is co-ordinated by EUROSTAT. We have selected families in which both spouses are between 20 and 64 years old and work as employees in five representative countries (France, Germany, Italy, Spain and UK). Those families lacking the required information have been excluded, resulting in a total sample ranging from 2,342 households in the UK to 4,360 in Italy. Households can stay in the survey all the periods or leave at any period, because of different reasons, as failing to meet the requirements or to provide some information. In this regard, the average period of permanence of a household in the survey varies from 2.4 years in Italy to 4.3 years in the UK.

Table 1 shows the name, and the average value of the main variables used in the analysis. The dependent variables are the number of weekly hours dedicated to paid work, whilst the right hand side variables include hourly wages, household non-labour income, the distribution factor and a variable indicating the number of children under 14 living with their parents. The hourly wage rates, as previously stated, are computed as the annual labour income divided by the number of weekly hours worked times 48 weeks in a year. To control for the endogeneity bias, hourly wage rates are regressed, separately for men and women, on a set of personal and labour characteristics. 11 The fitted wages are then incorporated to the estimation of the labour supply equations.

The annual household non-labour income includes both non-work private income and total social insurance receipts. 12 The distribution factor is computed as the ratio between 10 Since income variables refer to the period prior to the interview and the remaining data refer to the current period, the last year is lost for estimation. Thus, the time-series dimension reduces to seven years. 11 These include experience and squared experience, three dummy variables indicating educational attainment, nine occupational dummies, seven firm size dummies, one dummy for the public sector and one for fixed-term contracts. Results of these estimations are not shown but are available from the authors upon request.

12 All income variables have been deflated by the annual mean of the Consumer Price Index (CPI) base 1992 5). The intuition for this variable is that as children grow up they do not need attention from their parents. 13 Although the database provides information about the number of children and the hours devoted by the household to the care of children and/or adult members, there is no information about the time dedicated to housework. As a consequence, we have considered that all the time that an individual does not spend work is devoted to leisure. 14 In this way, we are forced to ignore the possibility of both household production (see Apps and Rees, 1997; Chiappori, 1997; Aronsson et al., 2001, for discussions on this matter) and public goods consumption (see, e.g., Fong and Zhang, 2001).

(Table 1)

With respect to the dependent variables, a major finding for the five country samples is that husbands dedicate more hours than wives to paid work. Germany is the sample where this difference is the greatest, 44.51 hours/week(h/w) for husbands as against 36.40 h/w for wives, whilst Italy exhibits the lowest, 40.04 h/w for husbands as against 34.05 h/w for wives. The UK is the country where both spouses spend the most time, as a whole, at paid work, whilst Italy shows the lowest hours per week.

Although hourly wages are expressed in euros, they are not comparable given that purchasing power parity is not the same across countries. This is also true for the family non-labour income. Within each country, male wages are clearly higher than female wages.

Spain is the country in which the fraction of the wife's non-labour income (over the household non-labour income) is the highest, at 80%; in France, Germany and Italy it is and are expressed in euros. The ECHP provides the non-labour income of each of the spouses and the total household income. We have computed household non-labour income as the difference between the total household income minus the total labour income. 13 Initially, we also introduced the age and squared age into the equation. However, these were always nonsignificant and were eliminated from the final estimation. Similarly, the average number of these children is in the range between 0.81 in France and 0.57 in Germany.

We simultaneously estimate the system of equations, and then perform the test to determine whether or not the collective model is adequate for describing the household labour supply behaviour in the sample countries. This consists of testing the restriction = , for the semilog, or the quadratic functional form, respectively.

IV. RESULTS AND POLICY IMPLICATIONS

For the five sample countries we estimate three models for the two parametric specifications, namely, the pool, the fixed effects and, finally, the random effects. Table 2 shows the LM test that allows us to choose between the pool estimation and the panel estimation, as well as the Hausman test that illustrates how to choose between the fixed and the random effects models. First, the LM test shows that the specification which considers individual effects is preferred to the pool estimation. This requires controlling for the unobserved individual heterogeneity present in the five sample countries. Additionally, the

Hausman tests reveal that unobserved individual heterogeneity may be correlated with some of the regressors. In these circumstances, the fixed effects model is preferred over the random effects model and, therefore, the results presented in our tables only correspond to this fixed effects estimation. 15 The top block of Table 2 shows the estimation for the semilog specification, whereas the bottom block shows those for the quadratic. Direct interpretation of the coefficients on wages and on non-labour income is not straightforward, given the non-15 Since this is the specification finally chosen, a time-varying distribution factor is selected to avoid the elimination of constant regressors by mean-differencing, as well as the collinearity arising from variables whose differences are constant over time, as, for example, gender differences in age or in experience, which initially were also considered as possible alternative distribution factors. linearities in the labour supply equations, in such a way that the discussion of these is left until later, when we interpret the elasticities.

However, before that, some general comments can be made. We first observe that collective modelling is non-rejected for both specifications in Italy; it is only rejected for the quadratic formulation in France, Spain and the UK; and rejected in both forms for Germany, although in this case the quadratic form closely approaches non-rejection. This appears in the last row for each block of estimates, where the t-statistic is shown for the restriction tested. 16 If we are ready to assume that the quadratic form is more appropriate than the semilog, since it allows for a more flexible specification, we can accept that a cooperative modelling of intrahousehold decisions accurately reflects reality in these five EU countries. Thus, in the following we focus only on the results obtained from such specification.

(Table 2)

Secondly, the sign of the household non-labour income, y, the distribution factor, s, and the number of children are clearly informative. Household non-labour income is significant in almost all cases. For wives, the coefficients are always negative, indicating that higher income is related to lower hours supplied. Since those coefficients are positive for husbands, save in the UK, they indicate that women are more sensitive than men to the family non-labour income in the total number of hours of paid-work supplied. This may be interpreted as women dedicating more time to housework and caregiving tasks, a phenomenon commonly found across countries. Only the UK couples show a different pattern of behaviour, perhaps reflecting a stronger desire to work fewer hours -as nonlabour income rises-given that they are the ones who jointly provide more hours of paidwork among the five countries analysed (see Table 1). Secondly, a higher fraction of the non-labour income that corresponds to the wife is associated with a lower number of hours supplied by the wife in all the countries, except in the UK. Again, this result may be 16 Note that the restrictions tested are if 3 With respect to the UK, note that this is a country in which the average share is considerably lower than in the others, such that a rise in this share may stimulate British women to engage in more hours of work. In Spain and France, a higher share is related to a higher number of hours supplied by the husband, probably indicating some type of substitutability or specialisation within the family. The contrary is the case in the other countries, which may reflect some form of income effect, by which a higher value in the share of wives non-labour income induces less work outside the home by the husbands.

Thirdly, when there are children under 14 in the household, the number of hours supplied by the wife is lower, whereas the impact on the husband's hours of work is non-significant in Italy, Germany and the UK, but clearly positive in Spain and France. In fact, in both these latter, the husband's labour supply increases, confirming the existence of a specialisation within the household, see Lundberg and Rose (2000).

As mentioned earlier, in order to analyse the effects of both wages and non-labour income on the hours of work supplied, it is appropriate to compute the elasticities, evaluated at the sample means, for each of the spouses. We first indicate that our values are in line with those obtained in the recent literature (Vermeulen, 2005). Since the collective model is generally non-rejected under the quadratic specification, we refer only to the results shown in the second column of Table 3 for each of the countries. A first result is that a rise in the wage of wives has no (or a reduced) effect on the hours supplied in paid-work by their husbands, labelled in Table 3 as Eh h w w , whereas it has a great influence on the hours supplied by the wives themselves, Eh w w w . In Germany and the UK, such influence is positive, and in France and Italy, negative. It can then be deduced that German and British wives are in the upward sloping section of their labour supply functions, whereas French and Italian wives are in the section in which the curves are backward bending. As regards a rise in the husband's wages, it has no impact on the hours supplied by their wives, except a strongly negative influence in the case of Spain, see the row headed by Eh w w h . In Spain, a rise in the husband's wages reduces the hours supplied by the husband, as it does in Italy, whereas it increases in the other countries -though, not in a significant way in France (see Eh h w h ). Similarly, as in the case of wives, Spanish and Italian husbands are in the Taken together, it seems that German and British couples react positively to rises in their wages, which may indicate a relatively lower valuation for leisure (or domestic work), such that a substitution effect dominates. By contrast, in the three Mediterranean countries, when the price of leisure rises, this is relatively more highly valued, such that the income effect dominates. Overall, these results may be interpreted as a distinct view of paid-work activities from the Northern perspective compared to the view from the Mediterranean.

When the rise in total income is only due to a higher availability of non-labour sources, a first clear result is that, unequivocally, wives always supply fewer hours of paid work, these results are shown in the row Eh w y. In the case of husbands the evidence is mixed: French and Spanish men tend to supply more hours, the contrary being observed in the other three countries (Eh h y). Again, this shows that specialisation may be driving family behaviour in France and Spain, such that men are more willing to engage in paid work, whereas wives dedicate more time to other activities (more than likely, housework or caregiving). In the other countries, higher non-labour income leads to fewer hours of work supplied by the couple, showing a clear preference for those alternative activities.

(Table 3)

From these results some policy implications can immediately be derived. Our estimates indicate that female labour supply is strongly influenced by both the amount of family income and the presence of children in the household. In order to make women's decisions on labour supply less dependent on these "family" factors, more equality between genders must be the target in the process of policy design. This may require two lines of approach. First, a general campaign by which female work should be valued, not only as a source of income, but also as a way of self-satisfaction and fulfillment. In this context, spouses should be committed to the sharing of housework, including childcare, in such a way that the labour supply of wives becomes less dependent on those "family" factors. 17 Second, the implementation of some measures aiming to make paid-work and housework compatible. Specifically, flexible time schedules that allow childcare during working time, and the establishment and growth of kindergardens at the work place with flexible timetables.

V. SUMMARY AND CONCLUSIONS

The aim of this paper has been to analyse the intra-family distribution of paid-work time in five representative EU countries (France, Germany, Italy, Spain and the UK). To that end,

we have formulated two labour-supply specifications, semilogarithmic and quadratic, which have been estimated for the five national samples by using the European Community

Household Panel-ECHP (1994-2001).

After proving that, in all cases, fixed effects is the preferred estimation method, we have observed that the collective model is non-rejected for the quadratic specifications in all sample countries, save Germany. With respect to estimations, we can draw the following conclusions. First, as a whole, working wives in Spain, the UK and Germany, represent the portion of the labour supply in which a wage increase leads to the supply of more hours of work. Conversely, in Italy and in France, wage increases lead women to devote more time to activities different from paid-work. Secondly, in all countries non-labour income is a significant variable in the amount of work supplied by wives: when this increases, labour supply decreases. Thirdly, the share of the wife's non-labour income over the household non-labour income is a relevant variable in deciding the amount of working hours supplied by each spouse. It has a negative effect in the case of wives, and a positive one in the case of husbands (except in the UK, where this share is considerably lower than in the rest of the sample countries). Fourthly, children under 14, who require more time, have a negative effect on the labour supply of wives, but not on that of husbands. Overall, it seems that, with some minor differences across countries, wives' labour supply is more affected by children and the proportion of non-labour income than that of men. Similarly, the response In short, we have noted that there is homogeneous performance across countries with respect to the behaviour of wives, with their labour supply being more affected than that of husbands by child care and non-labour income. Furthermore, the direction of the

  paid-work, and how these hours are shared out among the family.

  this restriction is empirically satisfied, we can obtain a characterisation of the intrahousehold allocation process in terms of the following expression for the sharing rule:

  s non-labour income and the total non-labour income of the household. As regards the sociodemographic variables, most of these are included in the wage regression and then only one indicating the number of children under 14 living in the household is considered in the estimation of the systems in (

  showing more flexibility than men in reducing the number of worked hours.

  of their labour supply curves, whereas German and British husbands are in the upward sloping section.

  17 A recent study byCouprie (2003) shows that today, in the UK, the younger generation does not appear to

  have experienced a change of attitude regarding gender equality in the family. by women is clearly dependent on own hourly wage, household non-labour income, the factors that contribute to the household non-labour income and the number of children. However, there is no clear evidence across countries of the influence of her spouse's wage on the number of paid-work hours.

=

  . Finally, s is the wife percentage of total nonlabor income.

Page 5 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

	B
	F o
	r
	P
	e e r
	R e v i e w
	6

  ).In order to be general enough to capture different behaviour across countries, we also use a quadratic specification. This offers several other advantages, in that, in addition to being linear in parameters, it provides a greater degree of flexibility since it includes the quadratic terms in wages, thereby allowing for backward bending labour supply. It operates

														Submitted Manuscript	Page 10 of 28
	in the following way:
	h	A	0 = +	1	A2	+	2	B 2	+	3	A B	+	4	A	+	5	B	+	6	y	7 + s + z 8
	h	B	0 = +	1	A2	+	2	B 2	+	3	A B	+	4	A	+	5	B	+	6	y	7 + s + z 8
							F o
										r
												P
													e e r
																		R e v i e w
	7 Whenever 4	4 / is different from 5 5 /
														11

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

12 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

															Submitted Manuscript
	h	A	0 = +	1	log		A	+	2	log	B	+	3	log		A	log		B	+		4	y	+			5	s	+	6	z	+	A	(5)
	h	B	0 = +	1	log		A	+	2	log	B	+	3	log		A		log		B	+		4	y	+			5	s	+	6	z	+	A
	h	A	0 = +	1	A2	+	2	B 2	+	3	A B	+	4	A	+	5	B	+	6	y	+	7	s	+	8	z	+	A
	h	B	0 = +	1	A2	+	2	B 2	+	3	A B	+	4	A	+	5	B	+	6	y	+	7	s	+	8	z	+	B
	where		(	)
					F o
										r
												P
														e e r
																				R e v i e w
															13

Page 13 of 28 Editorial

	Submitted Manuscript
	III.2. Data
	The data used in this work comes from the panel structure which results from the eight
	waves of the EHCP
	F o
	r
	P
	e e r
	R e v i e w
	14

Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

  14 Considering only the couples that offer information about hours caring for children or adults severely reduces the samples. whereas in the UK it is less than 40%. As regards children, the proportion of households with children under 14 varies from 35.2% in Germany to 51.7% in France.

	over 60%,
	F o
	r
	P
	e e r
	R e v i e w
	16

Page 16 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

	Submitted Manuscript
	F o
	r
	P
	e e r
	R e v i e w
	17

Page 25 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

  influence of own wages, household non-labour income, the proportion of non-labour income coming from wives, and the number of children, is the same across countries. As regards husbands, however, the results are quite different, the only common feature being the reduced influence of both the number of children and the household non-labour income in deciding the number of hours devoted to paid-work.

	Submitted Manuscript
	Vermeulen, F. (2005) And the winner is... An empirical evaluation of unitary and collective
	labour supply models, Empirical Economics, 30, 711-734.
	Weiss, Y., Willis, R.J. (1985) Children as collective goods and divorce settlements, Journal
	of Labour Economics, 19, 268-292.
	Wooldridge, J.M. (2002) Econometric analysis of cross section and panel data, MIT Press.
	Yamada, T. and Yamada T. (1987) Labor employment of married women in Japan: part-
	time work vs. full-time work, Eastern Economic Journal, 13, 41-48.
	F o F o F o F o F o
	r r r r r
	P P P P P
	e e r e e r e e r e e r e e r
	R e v i e w R e v i e w R e v i e w R e v i e w R e v i e w
	22 25 26

Table 1 .

 1 Descriptive analysis. Pooled data 1994-2001 Weekly hours. b Hourly wages expressed in euros. c Expressed in euros.

	France 43.06 35.69 9.02 6.68 2,639.01 0.69 0.81 Notes: a Page Germany Italy Spain UK Hours of work h a 44.51 40.04 42.83 45.66 Hours of work w a 36.40 34.05 36.70 38.43 Wage per hour h b 8.08 7.39 6.08 8.30 Wage per hour w b 6.29 6.29 4.60 6.87 Family nonlabor income, y c 2,217.37 2,238.38 1,307.18 2,041.87 % nonlabor wife income, s 0.68 0.62 0.80 0.36 Children (<14) 0.57 0.70 0.72 0.62

26 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

	Submitted Manuscript
	F o
	r
	P
	e e r
	R e v i e w
	27

Table 2 .

 2 Parameter and test results. Fixed effects model -statistics within parentheses. LM tests pool estimation against panel estimation. Hausman tests fixed effects against random effects. t-ratio tests the hypotheses of the collective model in the semilog form,

		France	Germany	Italy	Spain	UK
		W	H	W	H	W	H	W	H	W	H
	Semilog
	logw w	-0.025	0.358	0.800	-0.341	0.014	0.209	-0.177	0.576	1.901	0.913
		(-0.44)	(3.62)	(12.54)	(-3.06)	(0.10)	(1.07)	(-3.73)	(7.92)	(16.39)	(4.05)
	logw h	0.046	0.056	-0.613	0.637	0.446	-0.046	-0.683	-0.380	-0.609	0.678
		(0.56)	(0.74)	(-7.37)	(7.68)	(2.00)	(-0.24)	(-8.95)	(-7.63)	(-4.25)	(3.99)
	logw h logw w -0.071	-0.072	0.220	0.220	-0.074	-0.074	0.084	0.084	0.644	0.644
		(-1.90)	(-1.90)	(5.99)	(5.99)	(-0.77)	(-0.77)	(6.46)	(6.46)	(7.83)	(7.83)
	y	-0.825	1.027	-0.897	-1.708	-1.223	-1.054	-0.097	0.028	-5.386	-6.224
		(-4.96)	(6.04)	(-2.02)	(-4.00)	(-2.45)	(-2.11)	(-5.27)	(1.56)	(-3.23)	(-3.8)
	s	-2.444	2.944	-1.724	-0.861	-0.826	-0.375	-4.315	2.131	1.734	-0.867
		(-11.42)	(14.01)	(-5.21)	(-2.81)	(-1.99)	(-0.91)	(-12.13)	(6.30)	(5.11)	(-2.6)
	Children	-1.312	0.645	-1.551	0.133	-1.101	-0.118	-1.311	0.314	-2.728	0.142
	(<14)	(-12.98)	(6.72)	(-10.75)	(0.98)	(-3.58)	(-0.39)	(-7.37)	(1.96)	(-16.86)	(0.96)
	LM	18393	23729	4199	3766	16808
		(0.000)	(0.0000)	(0.0000)	(0.0000)	(0.0000)
	Hausman	97.78	508.63	47.38	215.23	272.23
		(0.000)	(0.0000)	(0.0000)	(0.0000)	(0.0000)
	t-ratio	15.54	3.70	-0.45	7.25	3.50
	quadratic
	w 2 w	-0.0002	-0.0002	-0.017	-0.0003	-0.006	-0.006	1.94 10 -6	9.32 10 -7	-0.037	-0.018
		(-2.95)	(-2.38)	(-9.64)	(-0.16)	(-1.54)	(-1.54)	(-6.06)	(2.74)	(-13.10)	(-
											3.12)
	w 2 h	0.0001	0.0002	0.005	0.005	0.001	-0.005	1.20 10 -7	1.26 10 -6	-0.004	0.004
		(0.43)	(1.04)	(2.28)	(3.56)	(0.32)	(-1.33)	(0.57)	(7.45)	(-1.14)	(0.25)
	w w w h	-0.0008	-0.0001	0.002	-0.095	-0.005	-0.0005	1.79 10 -6	-1.64 10 -6	-0.005	0.001
		(-0.43)	(-0.55)	(1.38)	(-2.38)	(-1.38)	(-0.15)	(6.22)	(-5.86)	(-1.48)	(0.30)
	w w	-0.001	0.041	0.530	0.217	0.123	0.046	0.001	0.0005	1.392	0.261
		(-0.14)	(4.26)	(13.54)	(4.93)	(1.53)	(0.57)	(1.92)	(0.88)	(20.44)	(2.95)
	w h	-0.006	-0.004	-0.172	-1.785	0.087	-1.071	-0.004	-0.004	0.024	0.084
		(-0.34)	(-0.28)	(-3.87)	(-4.18)	(1.09)	(-2.14)	(-6.48)	(-8.46)	(0.34)	(1.70)
	y	-0.786	0.990	-0.895	0.140	-1.241	0.120	-0.094	0.032	-4.737	-6.525
		(-4.72)	(5.81)	(-2.01)	(1.03)	(-2.49)	(1.54)	(-5.12)	(1.73)	(-2.84)	(-4.0)
	s	-2.390	2.867	-2.457	-0.817	-0.827	-0.413	-4.285	2.315	1.552	-0.870
		(-11.10)	(13.57)	(-5.44)	(2.66)	(-2.00)	(-1.00)	(-11.76)	(6.75)	(4.56)	(-2.6)
	Children	-1.927	0.638	-1.523	0.140	-1.088	-0.148	-1.668	0.344	-2.666	0.113
	(<14)	(-14.50)	(6.62)	(-10.54)	(1.03)	(-3.54)	(-0.49)	(-9.21)	(2.12)	(-16.43)	(0.76)
	N	11555	12795	10465	8983	10069
	LM	9591	23461	4264	3565	16443
		(0.000)	(0.000)	(0.000)	(0.0000)	(0.000)
	Hausman	382.43	542.72	47.16	68.00	223.73
		(0.000)	(0.000)	(0.000)	(0.0000)	(0.000)
	t-ratio	0.69	2.61	0.12	1.70	-0.19
	Notes: t

Page 27 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

	Submitted Manuscript
	F o
	r
	P
	e e r
	R e v i e w
	28

Table 3 .

 3 Elasticities

		France	Germany	Italy	Spain	UK
		Sem	Quad	Sem	Quad	Sem	Quad	Sem	Quad	Sem	Quad
	Eh w w w	-0.009	-0.033	0.039	0.048	-0.005	-0.032	0.011	0.001	0.075	0.119
		(-2.22)	(-2.88)	(16.21)	(8.89)	(-0.81)	(-2.25)	(5.31)	(0.17)	(22.53)	(22.9)
	Eh h w h	-0.005	0.026	0.027	0.024	-0.006	-0.041	0.004	-0.065	0.036	0.012
		(-2.07)	(1.35)	(18.52)	(2.77)	(-1.53)	(-2.47)	(2.81)	(-11.41)	(14.07)	(2.88)
	Eh w w h	-0.006	0.005	-0.002	-0.001	0.008	0.029	-0.003	-0.055	0.010	-0.007
		(-1.90)	(0.20)	(-0.77)	(-0.11)	(1.47)	(1.51)	(-2.36)	(-7.73)	(2.82)	(-1.49)
	Eh h w w	0.002	0.022	0.006	-0.008	0.0003	-0.010	0.0027	0.005	0.044	0.010
		(0.37)	(2.34)	(1.87)	(-1.75)	(0.04)	(-0.83)	(10.94)	(1.04)	(7.89)	(2.01)
	Eh w y	-0.008	-0.007	-0.002	-0.002	-0.001	-0.001	-0.006	-0.006	-0.004	-0.003
		(-4.96)	(-4.72)	(-2.02)	(-2.01)	(-2.45)	(-2.49)	(-5.27)	(-5.12)	(-3.23)	(-2.84)
	Eh h y	0.008	0.008	-0.003	-0.003	-0.0009	-0.0009	0.001	0.002	-0.004	-0.003
		(6.04)	(5.81)	(-4.00)	(-4.18)	(-2.11)	(-2.14)	(1.56)	(1.73)	(-3.82)	(-4.00)
	Note: t-statistic within parentheses
											Deleted: ¶
											Section Break (Next Page)

Page 28 of 28 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

Habitually, only the total household consumption is observed (and not the individual consumptions). We then talk of the Hicksian aggregate commodity.

According to this view, the household optimum allocation can vary as a consequence of a change in the nonlabour income both, directly, via the usual income effect, and indirectly through a shift in the bargaining

Empirical evidence on household labour supply for a number of countries has been found in, for example,[START_REF] Kawaguchi | Testing neoclassical models of household labour supply[END_REF][START_REF] Pradhan | Household labor supply in urban areas of Bolivia[END_REF][START_REF] Seaton | Neoclassical and collective rationality in household labour supply[END_REF][START_REF] García | Household labour supply with rationing in Spain[END_REF][START_REF] Barmby | Household labour supply in Britain and Denmark: some interpretations using a model of Pareto optimal behaviour[END_REF][START_REF] Flood | Household labor supply and welfare participation in Sweden[END_REF][START_REF] Doiron | Demands for child care and household labour supply in Australia[END_REF][START_REF] Iyigun | Endogenous gender power, household labor supply and the demographich transition[END_REF]. Additionally, particular studies on full-time/part-time work of spouses are, for example,Yamada and Yamada, 1987;[START_REF] Powel | Part-time versus full-time work and child care costs: evidence for married mothers[END_REF][START_REF] Connelly | Marital status and full-time/part-time work status in child care choices[END_REF]

ACKNOWLEDGEMENTS

This paper was partially written while José Alberto Molina was Visiting Fellow at the Department of Economics of the University of Warwick (UK), to which he would like to express his thanks for the hospitality and facilities provided. An earlier version of this paper has been presented at the 31 st Spanish Economic Analysis Meeting-2006 (Oviedo, Spain), with all the comments made by the participants being highly appreciated. Finally, the authors also thanks for the financial support provided by the Spanish Government (Grant SEC2005-06522), as well as by the Fundación BBVA. The usual disclaimer applies


Michael Overesch
email: overesch@zew.de.

Georg Wamser
email: wamser@ifo.de.

Corporate Tax Planning and Thin-Capitalization Rules: Evidence from a Quasi-Experiment

Keywords: Corporate Income Tax, Multinationals, Thin-Capitalization Rule, Differencein-Differences Estimation, Firm-Level Data JEL Classification: G32, H25, H26

This paper investigates tax-planning behaviour by means of inter-company finance and the effectiveness of government countermeasures via thin-capitalization rules. A simple theoretical model which considers the financing decision of a multinational company is used to obtain empirical implications. The empirical analysis, based on German inbound investment data from 1996 to 2004, confirms a significant impact of tax-rate differentials on the use of inter-company debt. The effectiveness of the German thincapitalization rule is tested by using legal amendments as natural experiments. The results suggest that thin-capitalization rules induce significantly lower internal borrowing. Hence, tax planning via internal finance is effectively limited by thin-capitalization rules.

Introduction

It is a well-established result, both theoretically and empirically, that taxes play an important role in determining the capital structure of companies (e.g. Modigliani andMiller, 1958, 1963;[START_REF] Desai | A multinational perspective on capital structure choice and internal capital markets[END_REF]. A multinational company in particular can choose its capital structure according to differences in international taxation, in order to minimize the tax burden of the whole company group. Borrowing from affiliates located in low-tax countries and lending to affiliates in high-tax locations will allow a deduction of interest payments from profits at high-tax locations and a reduction of the overall tax payments. Typically, high-tax countries attempt to restrict the use of inter-company loans by imposing so-called thin-capitalization or earning stripping rules to limit adverse revenue consequences. In recent years, increased attention has been given to the role of multinationals' profit shifting via inter-company loans. For example, the number of EU member countries that rely on some form of debt-to-equity restriction has increased from 8 in 1996 to 16 in 2005. From a theoretical perspective, these rules are suitable for limiting profit shifting (e.g. [START_REF] Fuest | Corporate tax policy, foreign firm ownership and thin capitalization[END_REF][START_REF] Panteghini | The capital structure of multinational companies under tax competition[END_REF]. Nevertheless, empirical evidence as to whether governments have been successful is still rare. Indeed, this is crucial for policy makers who want to defend tax revenues against multinationals' crosscountry tax planning. However, effective thin-capitalization rules possibly imply less financial flexibility for some firms and adversely affect investment. In order to evaluate this trade-off, it is important to find out more about the impact of thin-capitalization rules on financial decisions and the associated tax-revenue effects.

In providing evidence on the tax sensitivity of companies' capital structure choices, previous empirical studies have usually not taken into account thin-capitalization rules. However, for a sample of US controlled affiliates, [START_REF] Desai | A multinational perspective on capital structure choice and internal capital markets[END_REF] show that higher local tax 2 rates are associated with higher debt-to-asset ratios. Their analysis points out that particularly internal borrowing of US companies reacts sensitively to taxation. This result is confirmed by Buettner et al. (2006a), and by [START_REF] Mintz | Taxation and the financial structure of German outbound FDI[END_REF] for German multinationals. So far, evidence for the effects of thin-capitalization rules on companies' decisions has only been provided by [START_REF] Buettner | The impact of thin capitalization rules on multinationals' financing and investment decisions[END_REF]. They find that thincapitalization rules effectively restrict debt finance but also affect investment levels of German outbound investments.

The question we address in this paper is whether thin-capitalization rules effectively restrict multinationals tax-planning behaviour. For our empirical research we take German inbound investment data. We expect German multinationals to engage in tax-planning activities, because hardly any other country has higher statutory corporate tax rates than Germany. We exploit legal amendments of the German thin-capitalization rule in 2001 and 2004, where only some legal forms were treated. This quasi-experimental setting enables us to use a difference-in-differences approach to identify whether thin-capitalization rules are successfully imposed. Our empirical analysis shows that both tax-rate differentials and thin-capitalization rules are crucial for multinationals' capital structures. In particular, our findings indicate that some companies, which were affected by a stricter thin-capitalization rule, subsequently adjusted their capital structure. Hence, governments in high-tax countries are, to some extent, able to restrict multinationals' profit shifting.

The paper is organized as follows. We begin with some institutional details about the German thin-capitalization rule. Thereafter, we set up a theoretical model, which considers the financing decision of a multinational and takes into account a thin-capitalization rule. In sections 4 and 5 we present the empirical investigation approach and the data. The empirical results are presented in section 6. Section 7 briefly concludes. It is reasonable to assume that a multinational company allocates inter-company loans optimally with respect to differences in international taxation. Hence, taxable profits are reduced by means of interest deduction. As a result, jurisdictions lose corporate tax revenue and consequently try to defend their tax base by imposing some form of restriction, for example thin-capitalization rules. These rules, such as section 8(a) of the German corporate income tax law (KStG), typically limit interest deduction up to a fixed relation between equity and inter-company debt, i.e. the interest paid for an excess leverage cannot be deducted from the tax base.

Empirical identification of thin-capitalization rules, however, is difficult. In practice its application depends on several additional aspects besides the pure debt-to-equity ratio.

Some inter-company loans are excluded, such as debt which fulfils the arm's length principle requirements or trade accounts payable due to internal deliveries of input goods. Furthermore, the debt-to-equity ratio is not only limited to internal leverage. For instance, a strategy called back-to-back finance, in which external debt is borrowed by an affiliate and simultaneously secured by a deposit of the parent company, is also prohibited by the German thin-capitalization rule. Whether back-to-back finance is ultimately considered as parent-company debt finance is often a matter of negotiation with tax authorities. Hence, we are not able to identify single companies in the data which are affected by thin- capitalization rules, although we know the exact German threshold levels. Therefore, we use changes in the rules for some company groups, depending on their legal form, to obtain exogenous variation in this crucial explanatory variable.

The German thin-capitalization rule only applies to foreign affiliates that are incorporated. 1

First-tier foreign partnerships are not affected by the restriction and constitute a suitable non-treatment group when looking at legal amendments of this rule as natural experiments.

The rule classifies two different types of incorporated companies. The first group of companies comprises ordinary corporations which are not classified as holdings. For purposes of this law, a holding is defined as a firm where more than 75% of total assets consist of shares in other corporations. For ordinary corporations, the allowed debt-to-equity ratio, called safe haven, was accepted at 3:1 before 2001. Yet, the safe haven debt-to-equity ratio was 9:1 in the case of a holding corporation, i.e. holdings could be used as loopholes. In 2001 and 2004, two important amendments of the German thin-capitalization rule were introduced. In 2001 the allowed debt-to-equity ratios were significantly reduced to 1.5:1 in the case of an ordinary corporation and to 3:1 in the case of a holding corporation, respectively. Nevertheless, a possible loophole in the shape of holding corporations remained. In 2004, this special rule for holding corporations was also abolished, i.e. the safe haven was generally constituted at 1.5:1 for every corporation. However, first-tier partnerships were not affected by these amendments.

To illustrate the change in the thin-capitalization rule, let us consider the following example. A German holding corporation has used internal debt as a source of finance, with an internal-debt-to-equity ratio of 4:1 in 2000. Hence, this corporation was allowed to deduct interest payments without restriction. Following the 2001 tax reform, however, the ac-1 One exemption would be the German rule introduced in 2004 which applies to cases in which a second-tier partnership is held by incorporated foreign affiliates. However, these cases are not considered in our analysis. cepted ratio was reduced to 3:1. Therefore, the corporation was no longer able to deduct interest payments for 5 percentage points of its debt-to-equity share. Nevertheless, the corporation was not prohibited from maintaining its debt-to-equity ratio. It is important to bear in mind, however, that not every corporationIirrespective of its legal formIis affected by a stricter thin-capitalization rule. It is highly possible that many firms fall below the accepted debt-to-equity relation anyway. We will come back to this point in section 6, discussing which implications this has for the estimated coefficients.

A Model

We explain the impact of company taxation on the choice of debt or equity as a subsidiary's source of finance by the following simple model. We consider a firm with two separate companies, of which the parent company is denoted by 1 and the subsidiary by 2. If the parent company and the subsidiary are completely financed with equity, profits are determined by

) ( ) 1 )( ( ) 1 )( ( 2 1 2 2 1 1 k k r t k f t k f + + = .
1 k and 2 k denote invested capital, ) ( j k f output at location j , 1 t and 2 t the statutory tax rates at the respective locations. The opportunity costs of own capital are ) (

2 1 k k r +
. Note that an exemption system of repatriated foreign profits is assumed.2

The parent company can decide on the type of capital that is used to finance the subsidiary.

Instead of equity, the parent company can provide capital by means of an inter-company loan. We denote the share of capital that is financed by such an internal credit as 2 Moreover, we assume that inter-company loans are associated with additional costs, for example agency costs, because of asymmetric information (see Jensen and Meckling, 1976;[START_REF] Myers | Determinants of corporate borrowing[END_REF]. Nevertheless, there are also non-tax reasons to use inter-company debt, such as short-term cash management between parent and affiliate, or the opportunity to control the local management through fixed annual interest payments (see [START_REF] Jensen | Agency costs of free cash flow, corporate finance and takeovers[END_REF]. Subsequently, the profit function of the firm can be described as

. )] ( ) ( ) ( [ ) ( ) 1 )( ( ) 1 )( ( 2 2 2 2 2 1 2 2 2 2 1 2 2 1 1 k g c t t i k k r t k f t k f i µ µ µ + + + + = (1)
Obviously, the transition of equity into inter-company debt implies a direct profit shift from the borrowing affiliate 2 to the lending parent company 1 if the tax rate of the borrowing affiliate is higher than the tax rate of the lender. The tax-rate differential between both locations creates incentives to use equity refinanced internal debt as a source of finance. It implies that profits are shifted to the lending affiliate. However, jurisdictions attempt to counteract these activities by imposing thincapitalization rules, which typically limit interest deduction. Hence, interest paid for an excess leverage cannot be deducted from the tax base if j µ is above a certain fixed j µ .

We now assume that country 2 introduces such a rule, denoted by 2 , where 1 2 = if the rule is effectively binding, and 0 otherwise. Whether the rule is binding for an affiliate depends on the difference between the actual inter-company debt share and the maximum accepted, i.e. whether a company is above the threshold level or not, and how the type of internal debt is classified for tax purposes. Note that 2 is always 0 if 2 2 µ µ < . We extend the profit function by the additional tax payments arising from excess leverage above the thin-capitalization rule limit:

2 2 2 2 2 2 ) ( k t i i µ µ . The profit function (1) becomes . )] ( ) ( ) ( ) ( [ ) ( ) 1 )( ( ) 1 )( ( 2 2 2 2 2 2 2 2 2 2 1 2 2 2 2 1 2 2 1 1 k g c t i t t i k k r t k f t k f i i µ µ µ µ µ + + + + = (2)
The optimal share of inter-company debt of subsidiary 2 financed by parent equity is obtained by the following first-order condition 0

) ( ) ( ) ( ! 2 , 2 2 , 2 2 2 1 2 2 = + µ µ µ µ c g t t t i i i
.

(3)

First, we consider the case without application of a thin-capitalization rule. Accordingly, the share of inter-company debt is determined by the tax-rate difference between the borrowing affiliate and the parent company, ) (

1 2 t t i
. The internal lending rate can be used to leverage the tax effect. However, the interest rate is not likely to be an important degree of freedom, as the arm's length principle easily applies.

Secondly, we consider cases where the thin-capitalization rule is applied. If 2 becomes 1, any tax incentive to use internal debt is effectively stopped. In this case, only the tax-rate level at the location of the lending parent has an impact. Additionally, inter-company debt used for non-tax reasons becomes more expensive, because inter-company interest payments are taxed twice. Assuming a binding case, this is an incentive to reduce 2 µ in order to avoid enforcement of the thin-capitalization rule.

We can derive comparative static properties by differentiating the first-order condition:

2 2 2 1 2 2 2 2 2 1 2 ] [ ] [ di t t t dt i i dt i i i i + + = . )] ( ) ( [ 2 2 , 2 2 , 2 µ µ µ µµ µµ d g c
First, let us consider the effect of an increasing tax rate at the lending company's location on the share of inter-company debt used by its affiliate. 5 The derivative equals

0 ) ( ) ( 2 , 2 2 , 2 2 1 2 < = µ µ µ µµ µµ g c i dt d . (4)
Expression (4) is always negative, i.e. the inter-company debt used by an affiliate decreases with an increasing tax rate at the parent's location. With regard to the marginal effect of an increasing tax rate at the affiliate's location we obtain

. 0 ) ( ) ( 2 , 2 2 , 2 2 2 2 2 2 = µ µ µ µµ µµ g c i i dt d i (5)
This expression is positive when the thin-capitalization rule is not enforced, i.e.

. 0 2 = Otherwise, if the share of inter-company debt is above the limits, i.e. 1 2 = , a tax 5 We assume a zero marginal tax-rate effect on the internal interest rate 2 i . This is a reasonable assumption, because the arm's length principle is easily applied to interest rates. 9 rate increase has no effect on the optimal share of inter-company debt. This can be explained by the fact that interest payments for the excessive debt cannot be deducted for tax purposes. Higher internal borrowing must be due to non-tax reasons, e.g. short-term cash management. Accordingly, a tax-rate variation does not matter for the optimal intercompany debt in this case. To sum up, the following proposition can be set up:

Proposition 1: The optimal share of inter-company debt should increase with an increasing tax rate at the subsidiary's location and decrease with an increasing tax rate at the parent's location. Hence, if the share of internal debt is tax driven, the share of intercompany debt should increase with an increasing tax-rate difference, (

1 2 t t i
).

Finally, let us consider a reform of the thin-capitalization rule. Germany significantly reduced the safe haven in 2001 and 2004, respectively. Given any distribution of µ , a stricter thin-capitalization rule is associated with more companies above the lower threshold level. Reductions of 2 µ increase the number of cases for which 2 becomes one, i.e.

where the rule is binding and enforced. However, it should be emphasized that identification of each affected company is difficult due to various reasons we already discussed in section 2. Equation (3) implies that tax incentives to use inter-company debt decrease. In this case, the firm reduces its debt share below the new threshold level and prefers equity as the marginal source of finance. Therefore, we would expect that the level of intercompany loans has, on average, decreased since 2001 and 2004, respectively.

Empirical Implications

We test the above propositions empirically by using data for German inbound FDI. A simple estimation approach to test proposition 1 could be a regression of the inter-company leverage, denoted by ICL, of an investment in country G (Germany) taken by firm k located in country j in period t on the tax-rate differential (

t j t G STR STR , ,
) and on some company-specific characteristics t j x , . Thus, a simple regression equation would be

ICL t j k t k t j k t j t G t j k a a x a STR STR a a ICL , , , , 2 , , 1 0 , , ) ( + +
+ + + = , (6)
where k a is a company-specific effect to control for heterogeneity between company groups. Furthermore, we control for German capital market constraints or aggregate shocks by a time-specific effect t a . Note that we are able to identify tax-rate effects because of cross-country and time variation in t j STR , . Following proposition 1, we expect a positive sign of the tax differential coefficient 1 a on ICL . For inter-company debt, which is refinanced by equity, the local interest rate at the lending parent country should be irrelevant.

Only the German lending rate could be of importance, as it is used as the arm's length benchmark by the German tax authority. However, we implicitly control for the German lending rate by time effects t a , since every inbound investor faces the same lending rate.

We cannot explicitly identify companies which are affected by stricter thin-capitalization rules. However, two reforms in 2001 and 2004 constitute exogenous sources of variation, unambiguously affecting the average value of our dependent variable for some groups.

ICL t j k t k post TGROUP TGROUP t j k t j t G t j k b b D D b D b x b STR STR b b ICL , , 4 3 , , 2 , , 1 0 , , ) ( + +
+ + + + + = . (7)
The treatment group consists of both ordinary corporations and holding corporations. It is important to bear in mind that partnerships were not treated by the German thincapitalization rule and constitute a suitable non-treatment group of the reform in 2001.

Group-independent time trends are absorbed by t b . Therefore, we implicitly control for yearly variations in German tax rates. The treatment effect is measured by 4 b , where we expect a negative sign.

Furthermore, for the 2004 reform effect, we propose the following equation

ICL t j k t k TGROUP TGROUP t j k t j t G t j k c c D D c D c x c STR STR c c ICL , , 2004 4 3 , , 2 , , 1 0 , , ) ( + + + + + + + = , (8)
where only holdings were treated by the reform of the German thin-capitalization rule in 2004. Therefore, all other companies, incorporated and non-incorporated, constitute the non-treatment group. We would also expect a negative sign of the treatment effect 4 c . To sum up, the legal reforms enable us to test whether a thin-capitalization rule is effectively imposed. The tax reform of 2001 only hit ordinary incorporated as well as holding companies. Partnerships, however, were not affected. In 2004, only holding corporations were affected. We argue that groups are comparable, because we observe only affiliates of multinationals. Hence, we look at firms which have the same opportunities with respect to, for example, internal finance. Furthermore, we control for differences across single investments, for example by using the affiliate-specific turnover as a control variable. We additionally assume that there are no systematic changes in within-and between-group compositions. In fact, our group sizes are almost stable over time. To identify the effect on the treated companies, a further critical assumption is that both groups are equally affected by aggregate shocks.

Data and Descriptive Statistics

The empirical analysis uses the MiDi database for multinationals provided by the Deutsche Bundesbank. This is a comprehensive annual micro database of direct investment positions of German enterprises held abroad as well as of direct investment positions held in Germany by foreign companies. However, we employ only German inbound FDI data. The data provides information about the investment object's balance sheet, including further information on the type of investment and on the investor. A favourable characteristic of the data is the possibility to trace direct investment positions of individual firms over time.

The current version provides firm-level panel data for the period 1996 to 2004. The collec- According to equation ( 6), we calculate the inter-company loan, borrowed from the foreign parent company, to total capital ratios, ICL, using the Midi data. In order to control for company-specific variation in the accession to external debt, we employ the turnover as an indicator of the size and the affiliate's cash-flow in our regression analysis (see e.g. [START_REF] Panno | An empirical investigation on the determinants of capital structure: the UK and Italian experience[END_REF]. We expect a negative effect of a higher turnover when external and internal debt serve as substitutes (see Buettner et al., 2006a). As agency costs and the utility of intercompany debt may also vary across industries, we control for further heterogeneity by including dummies for 56 industries at the level of the affiliate.

F

6 Sec. 26 Aussenwirtschaftsgesetz (Law on Foreign Trade and Payments) in connection with Aussenwirtschaftsverordnung (Foreign Trade and Payment Regulations). Since 2002, FDI has to be reported if the participation is 10% or more and the balance sheet total of the foreign investment in Germany is above 3 million Euro. For details see [START_REF] Lipponer | Microdatabase Direct Investment -MiDi[END_REF]. Despite that previous years showed lower threshold levels, we apply this threshold level uniformly for all years in the panel. The statutory tax-rate differential constitutes the relevant measure to investigate the tax impact on the use of inter-company debt. The variable G STR contains German statutory profit tax rates; 7 foreign statutory tax rates are denoted by F STR . While using these two tax measures, bilateral tax-rate differences are constructed. Since the effective tax reduction from using debt might be zero if a subsidiary carries forward any losses for tax purposes (see [START_REF] Mackie-Mason | Do taxes affect corporate financing decisions[END_REF][START_REF] Francois | Tax loss carry-forwards and optimal leverage[END_REF], we include a dummy variable indicating whether some loss carry-forward is reported. Of course, the existence of some losses in the previous periods may capture other characteristics of the current decision problem of the company such as the expected performance of an affiliate. Thus, the overall effect on internal leverage is ambiguous. Table 2 displays basic information about the regression variables.

--Insert Table 2 about here --

Empirical Results

The empirical analysis involves panel-data regressions that include company fixed effects.

Hence, by using a within transformation, we generally control for all time-constant heterogeneity between company groups.

--Insert Table 3 about here --First of all, it is worth mentioning that all regressions show the expected tax rate effect. In 15 tax-rate differential between Germany and any other country is associated with a 1.9 percentage point higher internal-debt ratio. The magnitude of the tax effect is in line with earlier findings, see e.g. [START_REF] Desai | A multinational perspective on capital structure choice and internal capital markets[END_REF], or Buettner et al. (2006a). This result indicates that tax-rate differentials play an important role in the choice of financial structures. With regard to the effect of the German thin-capitalization reform in 2001, we control for systematic differences in the control and treatment group by introducing a treatment-group dummy variable, TGROUP D . First, we do not distinguish between different treatment groups in specifications ( 3)-( 5) of Table 3. While we interact the treatment group with a postreform dummy, POST D , for the whole period of 2001 until 2004 in column (3), we control for each single year in columns ( 4) and ( 5). In column (4) we observe that the treated group, i.e. holdings and incorporated firms, responds to the tighter thin-capitalization rule, and internal lending is reduced in the post-reform period. Column (5) shows that the reform was by no means anticipated. Rather, it took the companies one year to reduce their internal debt shares. This is possibly the result of a restructuring process which started in

2001.

A company is only affected if the thin-capitalization rule is binding, i.e. if the firm's debtto-equity share is above the maximally accepted share. In this case, the tax-planning firm should re-optimize its capital structure after the reform, if the construction so far has been optimal. Nevertheless, given the continuum of internal-debt-to-equity shares, not all corporations are affected. However, the mean share is reduced for the whole treatment group because of certain corporations. Consequently, the treatment effect would be much stronger if all treated corporations were noticeably affected.

--Insert 16

In Table 4 we split the sample into different treatment groups. As already mentioned, we have two treatment groups for the 2001 reform, i.e. holdings and ordinary incorporated companies. The results are generally comparable, apart from the fact that the number of observations is reduced. 8 The specifications (1)-( 3) consider ordinary corporations as the treatment group and partnerships as the control group. Specifications ( 4)-( 6) investigate the tax effects on inter-company debt of holding corporations; partnerships constitute the control group again. One major insight from Table 4 is that holdings adapt their capital structure much faster than ordinary incorporated companies. The restructuring process is possibly easier, and therefore faster for some reason. Moreover, the threshold level was reduced more severely for holding corporations. Another reason might be the potential role of holding corporations as special tax-planning entities.

The magnitude of the treatment effects can be interpreted as follows. For holding corporations, the 2001 thin-capitalization reform induced a decline in the share of internal debt borrowed from the foreign parent company of about 10 percentage points. This equals a reduction of approximately one-third, considering a pre-reform mean of 31.7 per cent intercompany debt to total capital.

Finally, we focus on the effect of the German thin-capitalization reform in 2004. This reform treated holding corporations only. The control group consists of all other legal forms.

The results in Table 5 show once more that some companies restructure, basically those which were affected by the stricter rule, and reduce their internal debt share, i.e. the mean debt share decreases for the treated group. Specifications (1) and ( 2) are based on observations from 1996-2004. However, regressions in columns (3) and ( 4) are only based on observations from 2001 to 2004, in order to avoid the effect of the first thin-capitalization 8 Note, however, that the group sizes are relatively constant for the 9 years in the sample. --Insert Table 5 about here --

Conclusion

We find that international tax-rate differentials play an important role in determining the share of internal lending to German affiliates. This confirms earlier results provided by other studies. The important insight of our empirical analysis is that German thincapitalization rules are effectively imposed. A reduction of the allowed debt-to-equity ratios, enforced by reforms in 2001 and 2004, respectively, induced significantly smaller shares of inter-company loans of incorporated companies.

Finally, we focus on revenue effects. Although we cannot estimate how much revenue Germany would lose in the absence of a thin-capitalization rule, rough estimations of the reform effects can be provided. First of all, we consider specifications (3) and ( 6) from .811 The dependent variable is the share of inter-company loans borrowed from the foreign parent company. Specifications (1) -( 3) are based on a sample of ordinary corporations and partnerships. Specifications (4) -( 6) are based on a sample, which consists of holding corporations and partnerships. Robust and clustered (country/year clusters) standard errors are in parentheses. A star denotes significance at 5% and two stars at the 1% level. All estimates include a full set of 5,105/738 firm and 56 industry fixed effects. .773 The dependent variable is the share of inter-company loans borrowed from the foreign parent company. Specifications (1) -( 2) are based on the whole sample, whereas (3) and ( 4) are based on a sample in which only observations from 2001 until 2004 are considered. Robust and clustered (country/year clusters) standard errors are in parentheses. A star denotes significance at 5% and two stars at the 1% level. All estimates include a full set of 5,257 or 3,196 firm, and 56 industry fixed effects.

Hence

Proposition 2 :

 2 If inter-company debt is a channel for shifting taxable profits, and thincapitalization rules can limit tax-planning behaviour, a reduction of the allowed debt-toequity ratio should lead to smaller shares of inter-company debt. Non-incorporated com-are not treated by the German thin-capitalization rule and its amendments, do not reduce their inter-company debt levels.

  test proposition 2 using a difference-in-differences approach (see for example[START_REF] Meyer | Natural and quasi-experiments in economics[END_REF].Different amendments of theGerman thin-capitalization rule have constituted three groups with different degrees of treatment. We use a dummy variable, TGROUP D , to denote the respective treatment group, i.e. the corporations which were treated by a specific amendment of the thin-capitalization rule in 2001 or 2004, respectively. Additionally, we use a dummy variable post D to indicate the post-amendment observations from 2001 to 2004. In this manner, an estimation equation for the 2001 reform effect can be described as

  . The magnitude of the 2004 reform effect is much lower compared to the 2001 reform. Here, the share of inter-company debt is reduced by approximately 4.2 percentage points. This equals a reduction of about one sixth, given the pre-reform level of 26 per cent inter-company debt to total capital.

Table 3 ,

 3

	F o r
	P
	e
	e r
	R e v
	i e
	w

specification (2) indicates, for example, that a 10 percentage point increase in the

7

Our measure takes into account that only half of all interest payments can be deducted from the tax base of the German trade income tax. Furthermore, we consider the country average of the local trade income tax.

  Table 4 about here --

	F o r
	P
	e
	e r
	R e v
	i e
	w

Table 4 .

 4 In 2002, the 2001 reform is associated with lower internal-debt-to-capital ratios of Secondly, using specification (4) of Table5, we estimate the additional tax revenue to be up to approximately 30 million Euros for the reform in 2004. Total amounts might be higher due to the treatment of indirectly held foreign affiliates or of other debt types. Moreover, the internal lending rate is not necessarily equivalent to the local lending rate. All estimated magnitudes can only be rough estimates, because we cannot take into account that multinationalsIin case of restrictions on their capital structure choiceIare able to shift profits through other channels, e.g. by transfer-price setting. The table shows a per-country apportionment of parent companies for German inward FDI from 1996 until 2004. Data are taken from the MiDi database provided by the Deutsche Bundesbank. Internal debt shares are country-specific mean values, determined by the level of balance sheet liabilities divided by total capital. The table also shows average statutory tax-rate differences between Germany and respective foreign locations, as well as the yearly average number of investments. a) Not reported due to data protection, because the number of observations is small.

	approximately 0.053 for ordinary corporations and 0.151 for holding corporations, respec-
	tively. Given the respective mean values, Germany was able to retain, on average, 71,700
	Euros per ordinary corporation and 1,807,000 Euros per holding corporation in additional
	tax revenue through the tightening of the thin-capitalization rule. 9 Considering the number
	of treated corporations, we estimate an amount of approximately 260 million Euros in ad-

Table 2 :

 2 Descriptive Statistics Observations: 19,379. Minimum and maximum values are averages of the 3 smallest (highest) values.

	Variable		Definition	Mean Std.Dev.	Min.	Max.
	ICL				internal debt share from	.280	.262	.001	.992
					foreign parent company
	G STR	STR	F	tax-rate difference	.003	.072	-.152	.281
	STR	F			foreign tax rate	.354	.071	.100	.532
	Loss carry-forward binary	.431	.495	0	1
	Turnover		turnover in €1,000	65,088	247,678	1,000 11,000,000

Table 3 :

 3 Inter Company Loans -Thin-Capitalization Reform in 2001 The dependent variable is the share of inter-company loans borrowed from the foreign parent company. Robust and clustered (country/year clusters) standard errors are in parentheses. A star denotes significance at 5% and two stars at the 1% level. All estimates include a full set of 5,257 firm and 56 industry fixed effects.

							(1)	(2)	(3)	(4)	(5)
	STR G	STR	F	.188 (.088) *	.192 (.088) *	.206 (.087) *	.207 (.087) **	** (.088) .214
	TGROUP D						.001	.002	.018
									(.019)	(.019)	(.020)
	TGROUP D	xD	POST			-.025	*
									(.010)
	TGROUP D	x 1997			-.006
									(.018)
	TGROUP D	x 1998			-.016
									(.019)
	TGROUP D	x 1999			-.025
									(.017)
	TGROUP D	x 2000			-.024
									(.018)
	TGROUP D	x 2001			-.004	-.021
									(.012)	(.020)
	TGROUP D	x 2002			-.036	*	-.052	*
									(.015)	(.023)
	TGROUP D	x 2003			-.032	*	-.049	*
									(.011)	(.020)
	D TGROUP	x 2004			-.038 (.016) *	* (.023) -.054
	1997						-.003	-.003	-.003	-.003	.003
							(.005)	(.005)	(.005)	(.005)	(.017)
	1998						-.002	-.002	-.002	-.002	.013
							(.006)	(.006)	(.006)	(.006)	(.017)
	1999						-.001	-.001	-.002	-.002	.022
							(.005)	(.005)	(.005)	(.005)	(.016)
	2000						-.000	-.000	-.000	-.000	.022
	2001						(.006) .002	(.006) .002	(.006) .025 **	(.006) .006	(.017) .022
							(.005)	(.005)	(.011)	(.012)	(.019)
	2002						-.011	-.010	.013	.023	.038
	2003						(.006) -.024 **	(.007) -.024 **	(.011) -.001	(.015) .006	(.021) .020
	2004						(.006) -.022 **	(.006) -.022 **	(.010) .001	(.011) .013	(.018) .028
	ln(Turnover)			(.007) -.016 **	(.007) -.015 **	(.007) -.015 **	(.016) -.015 **	(.021) -.015 **
							(.004)	(.004)	(.004)	(.004)	(.004)
	Loss carry-forward		.009	.009	.009	.009
								(.005)	(.005)	(.005)	(.005)
	Observations			19,379	19,379	19,379	19,379	19,379
	Adj.	R	2				.711	.711	.711	.711	.711

Table 4 :

 4 Inter Company Loans -Thin-Capitalization Reform in 2001

	G STR	STR	F	(1) .201	*	(2) .216	*	(3) .224	*	(4) .783	**	(5) .750	**	(6) .737	**
							(.089)	(.088)	(.090)	(.287)	(.288)	(.294)
	TGROUP D						.003		.019				.019	.100
									(.019)	(.020)			(.074)	(.084)
	TGROUP D	xD	POST			-.025	*					-.101	**
									(.010)					(.018)
	TGROUP D	x 1997					-.006				-.044
											(.018)				(.028)
	TGROUP D	x 1998					-.017				-.017
											(.019)				(.035)
	TGROUP D	x 1999					-.026				-.036
											(.017)				(.032)
	TGROUP D	x 2000					-.024				-.085	*
											(.018)				(.034)
	TGROUP D	x 2001					-.020				-.125	*
											(.020)				(.038)
	TGROUP D	x 2002					-.053
				x 2003					-.050
				x 2004					-.052
											(.023)				(.046)
	1997					-.002	-.002	.003		-.006	-.006	.003
							(.006)	(.006)	(.017)	(.014)	(.015)	(.019)
	1998					-.002	-.002	.013		.007	.006	.009
							(.006)	(.006)	(.017)	(.013)	(.014)	(.018)
	1999					-.002	-.002	.022		.019	.015	.020
							(.005)	(.005)	(.016)	(.013)	(.014)	(.018)
	2000					-.000	-.000	.022		.008	.000	.016
	2001					(.006) .003	(.006) .026 *	(.017) .022	(.015) .026	(.015) .040	(.019) .040
	2002					(.006) -.010	(.011) .013	(.019) .039	(.020) .034	(.020) .046 *	(.023) .051 *
	2003					(.007) -.024 **	(.011) -.002	(.021) .021	(.020) .006	(.021) .018	(.024) .031
	2004					(.006) -.019 **	(.011) .003	(.018) .028	(.017) .018	(.018) .029 **	(.021) .048
	ln(Turnover)			(.007) -.014 **	(.012) -.014 **	(.021) -.014 **	(.020) -.019 **	(.020) -.017 *	(.025) -.018 *
							(.004)	(.004)	(.004)	(.007)	(.007)	(.007)
	Loss carry-forward	.009		.009		.009		-.011	-.013	-.014
							(.006)	(.005)	(.005)	(.015)	(.014)	(.014)
	Observations			18,787	18,787	18,787	2,196	2,196	2,196
	Adj.	R	2				.714		.714		.714		.806	.810

* -.151 * (.023) (.041) TGROUP D * -.193 * (.020) (.042) TGROUP D * -.221 *

Table 5 :

 5 Inter Company Loans -Thin-Capitalization Reform in 2004

	(1)	(2)	(3)	(4)

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

This is true for most European countries and Canada. Furthermore, the effect of a credit system equals an exemption system if t 2 > t 1. . Otherwise, the affiliate's tax rate increases to t 1 , depending on the time of retention. However, the general incentives remain if foreign profits are not distributed immediately.

We assume as price for the internal credit a German lending rate for credits to the private sector of 9.7% in 2002 (IMF, 2005), and the 2002 statutory tax rate, 32.88%. Page 18 of 25 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Acknowledgments Page 19 of 25 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

We are grateful to the Deutsche Bundesbank for granting access to the FDI database. Financial support by the German Research Foundation (DFG) is gratefully acknowledged.

We are also indebted to Thiess Büttner, Ulrich Schreiber, and an anonymous referee for insightful comments. Errors remain the responsibility of the authors.

< H -Public Economics, H26 -Tax Evasion < H2 -Taxation, Subsidies, and Revenue < H -Public Economics

Data Sources and Definitions

Firm-level data are taken from the micro-level dataset of the Bundesbank, see [START_REF] Lipponer | Microdatabase Direct Investment -MiDi[END_REF] for an overview. The inter-company debt share from the foreign parent company is determined by the level of balance-sheet liabilities in the respective category divided by total capital consisting of registered capital, capital reserves and profit reserves, as well as internal and external debt.

Corporate taxation data are taken from the IBFD, and from tax surveys provided by the tax advisory companies Ernst&Young, PwC and KPMG.