Marc Deloof
email: marc.deloof@ua.ac.

Wouter Van Overfelt

Evidence from Pre-World War I Belgium

come

I. Introduction

For many firms, trade receivables are an important part of their assets. This raises the question why firms extend credit to their customers. It has been argued that trade credit is used to reduce the transaction costs of paying bills [START_REF] Schwartz | An economic model of trade credit[END_REF][START_REF] Ferris | A transactions theory of trade credit use[END_REF], to smooth demand and price discriminate [START_REF] Emery | An optimal financial response to variable demand[END_REF][START_REF] Brennan | Vendor financing[END_REF], or to allow customers to assess product quality before paying (e.g. [START_REF] Smith | Trade credit and informational asymmetry[END_REF][START_REF] Lee | Product risk, asymmetric information, and trade credit[END_REF][START_REF] Bastos | An agency model to explain trade credit policy and empirical evidence[END_REF]. Finance based trade credit models argue that suppliers have significant cost advantages over financial institutions in providing credit to their customers (e.g. [START_REF] Schwartz | An economic model of trade credit[END_REF][START_REF] Emery | A pure financial explanation for trade credit[END_REF][START_REF] Biais | Trade credit and credit rationing[END_REF]. Furthermore, suppliers desiring to maintain a long-term relationship with customers may grant more concessions to a customer who is in need for extra liquidity than would be granted lenders in a competitive credit market [START_REF] Wilner | The exploitation of relationships in financial distress: the case of trade credit[END_REF][START_REF] Cuñat | Trade credit: suppliers as debt collectors and insurance providers[END_REF]. In countries where formal financial institutions are underdeveloped, trade credit may be an important informal financial institution [START_REF] Mcmillan | Interfirm Relationships and Informal Credit in Vietnam[END_REF][START_REF] Demirgüç-Kunt | Firms as financial intermediaries: evidence from trade credit data[END_REF][START_REF] Fisman | Trade credit, financial intermediary development, and industry growth[END_REF][START_REF] Cull | Formal finance and trade credit during Chinas transition[END_REF]: firms with better access to credit channel short-term funds from financial institutions via trade credit to financially weaker firms.

In this paper, we go back to the start of the 20 th century to test the hypothesis that bank-affiliated firms with better access to credit redistributed funds via trade credit to financially weaker customers. Belgium before World War I provides a particularly interesting environment to test this hypothesis. Investors were poorly protected and faced severe information problems [START_REF] Théate | Bank information monopoly and trade credit: do only banks have information about small business[END_REF], while the banking sector was strongly developed (e.g. [START_REF] Rajan | The Great Reversals: The Politics of Financial Development in the 20th Century[END_REF]. A limited -3 -number of universal banks offered a wide range of services to firms (e.g. [START_REF] Durviaux | La Banque Mixte: Origine et Soutien de lExpansion Economique de la Belgique[END_REF][START_REF] Kurgan-Van Hentenryk | Finance and Financiers in Belgium, 1880-1914[END_REF], and many Belgian firms had close ties with these banks [START_REF] Van Overfelt | Do Universal Banks Create Value? Universal Bank Affiliation and Company Performance in Belgium[END_REF]. We investigate the determinants of trade credit granted by suppliers to their customers. Our findings, based a sample of 535 firm-year observations for 125 listed firms in the coalmining, tramways, railways and textile industries in the period 1905-1909, generally confirm the financing role of trade credit. We find that firms in which a bank had an equity stake redistributed capital via trade credit to other firms.

However, firms with a banker on their board but without a bank equity stake did not provide more credit to their customers. Older firms and firms with higher cash flows also provided more trade credit. Consistent with findings for contemporary developing countries, our results suggest that a century ago, trade credit was a tool for channelling funds from firms with close bank ties to other firms. As such, our study provides a unique contribution to the literature.

The remainder of the paper is structured as follows. Section 2 discusses the financing role of trade credit, while section 3 focuses on trade credit and bank relationships in pre-World War I Belgium. The sample and the construction of the variables is discusses in section 4. Regression results are presented in section 5. Section 6 concludes.

II. The financing role of trade credit

Finance based trade credit models argue that suppliers have significant cost advantages over financial institutions in providing credit to their customers. [START_REF] Petersen | Trade credit: theories and evidence[END_REF] -4 -advantage over outside financial institutions. Size and timing of the buyer's orders and his ability to take advantage of early payment discounts help the supplier in assessing the condition of the buyer's business and his creditworthiness. Second, the supplier has more control over the actions of the buyer if they have a repeated relationship. The supplier holds the threat of stopping the supply of inputs to the buyer if he has few alternative sources. The threat by a financial institution to withdraw future finance on the other hand may have little immediate impact on the borrower. Third, if the buyer defaults, the supplier may be able to seize the goods he supplied. Furthermore, suppliers desiring to maintain a long-term relationship with customers may grant more concessions to a customer who is in need for extra liquidity than would be granted lenders in a competitive credit market. Suppliers know that they may have to help customers in financial distress, so they will charge them a higher trade credit interest rate as compensation for providing insurance against liquidity shocks [START_REF] Wilner | The exploitation of relationships in financial distress: the case of trade credit[END_REF][START_REF] Cuñat | Trade credit: suppliers as debt collectors and insurance providers[END_REF]. Trade creditors may also alleviate the problems of bank information monopolies: when the bank raises the interest rate to extract a rent, the borrowing firm can substitute bank debt by trade debt [START_REF] Théate | Bank information monopoly and trade credit: do only banks have information about small business[END_REF]. [START_REF] Burkart | In-kind financing: a theory of trade credit[END_REF] and [START_REF] Antov | Trade credit financing: how expensive is it really?[END_REF] argue that when suppliers have significant cost advantages over financial institutions in providing credit to their customers, trade credit may not only be a substitute for institutional financing, but may actually increase the availability of institutional financing: trade credit use mitigates moral hazard problems and may be perceived as a favorable signal that is informative of the creditworthiness of the borrower.

Recently, a new literature has emerged which argues that in countries where formal financial institutions are underdeveloped, trade credit is an important -5 -informal financial institution. [START_REF] Mcmillan | Interfirm Relationships and Informal Credit in Vietnam[END_REF] find that Vietnamese firms are more likely to extend trade credit if it trusts its customer enough to offer credit. They extend more credit to customers who find it hard to locate an alternative supplier and customers belonging to network of similar firms. [START_REF] Fisman | Trade credit, financial intermediary development, and industry growth[END_REF] show that industries with higher dependence on trade credit financing have higher growth rates in countries with weaker financial institutions. In such countries, firms with better access to credit may redistribute capital via trade credit to financially weaker customers (e.g. [START_REF] Marotta | Does trade credit redistribution thwart monetary policy? Evidence from Italy[END_REF][START_REF] Demirgüç-Kunt | Firms as financial intermediaries: evidence from trade credit data[END_REF][START_REF] Nilsen | Trade credit and the bank lending channel[END_REF]. These firms will act as "agents" for the financial institutions, channelling short-term funds from the financial institutions to their best use. Consistent with this view, [START_REF] Love | Trade credit and bank credit: evidence from recent financial crises[END_REF] find that after the emerging markets financial crises of the 1990s, firms with more vulnerable financial positions were more likely to reduce their supply of credit to customers. [START_REF] Cull | Formal finance and trade credit during Chinas transition[END_REF] find that state-owned Chinese firms, which have privileged access to loans, redistribute credit to private firms with less privileged access. Profitable private firms in China lend part of their formal credit via trade credit to their customers.

III. Bank relationships and trade credit in pre-World War I Belgium

As in many developing countries today, investors in pre-World War I Belgium were poorly protected and faced severe information problems. [START_REF] Théate | Bank information monopoly and trade credit: do only banks have information about small business[END_REF], a contemporary Belgian law scholar, argued that investor protection was much worse in Belgium than in other European countries such as Britain, France and Germany. In an environment characterized by high information asymmetries and agency costs, it can be expected that trade credit is an important financing tool (e.g. McMillan and Woodruff, 1999;[START_REF] Fisman | Trade credit, financial intermediary development, and industry growth[END_REF]. [START_REF] Miwa | Japanese industrial finance at the close of the 19 th century: trade credit and financial intermediation[END_REF] find that manufacturing firms in Japan at the turn of the last century indeed raised much of the money they needed through trade credit.

While investor protection was weak, banks played an important role in pre-World War I Belgium. The ratio of commercial and savings deposits over GDP indicates that the banking sector in 1913 was more important in Belgium (0.68) than in Germany (0.53), the U.S. (0.33), Japan (0.13) or the U.K. (0.10) [START_REF] Rajan | The Great Reversals: The Politics of Financial Development in the 20th Century[END_REF]. In the late 19 th century and the early 20 th century, banks in industrializing countries generally played a dominant role in corporate finance (e.g. [START_REF] Cameron | Banking in the Early Stages of Industrialization: A Study in Comparative Economic History[END_REF][START_REF] Carosso | Investment Banking in America: A History[END_REF][START_REF] Rajan | The Great Reversals: The Politics of Financial Development in the 20th Century[END_REF][START_REF] Fohlin | Finance Capitalism and Germanys Rise to Industrial Power: Corporate Finance, Governance, and Performance from the 1840s to the Present[END_REF].

The most prominent of these banks in Belgium were universal banks, which offered a wide range of services to firms: they took equity stakes, provided loans, assisted firms in the issuance of securities and were often involved in the management of affiliated firms (e.g. [START_REF] Durviaux | La Banque Mixte: Origine et Soutien de lExpansion Economique de la Belgique[END_REF][START_REF] Kurgan-Van Hentenryk | Finance and Financiers in Belgium, 1880-1914[END_REF].

These banks had close ties with many Belgian firms, either as a shareholder or through interlocking directorships [START_REF] Van Overfelt | Do Universal Banks Create Value? Universal Bank Affiliation and Company Performance in Belgium[END_REF]. Firms with close bank ties are likely to face less credit constraints than firms without bank ties.

Enhanced monitoring by the universal bank through close and continued interactions with the firm increases available credit for the firm. Delegated monitoring by a financial intermediary has lower monitoring costs than monitoring by a group of individual borrowers, and it avoids the free rider problem [START_REF] Diamond | Financial Intermediation and Delegated Monitoring[END_REF]. The Belgian universal banks indeed played an active role in the companies they controlled [START_REF] Kurgan-Van Hentenryk | The Société Générale[END_REF]. [START_REF] Carosso | Investment Banking in America: A History[END_REF][START_REF] De Long | Morgans Men Add Value? An Economists Perspective on Financial Capitalism[END_REF][START_REF] Ramirez | Did J. P. Morgans men add liquidity? Corporate investment, cash flow and financial structure at the turn of the Twentieth Century[END_REF][START_REF] Franks | The Origins of the German Corporation -Finance, Ownership and Control[END_REF]. A bank with access to more timely information and with incentives to produce more durable information can better certify offer prices in securities issues. The findings of [START_REF] Deloof | Who's Afraid of Universal Banks? Universal Bank Affiliations and Corporate Dividend Policy in Pre-World War I Belgium[END_REF] that in the period considered in this study, Belgian firms with a banker on their board paid higher and more stable dividends than firms without bank relationships is consistent with the "certification" role of bank directors.

A number of studies find that relationships with financial institutions indeed reduce credit constraints (see [START_REF] Ongena | Bank relationships: a review, in Performance of Financial Institutions[END_REF], for a review). Some of these studies focus on the early 20 th century. [START_REF] Ramirez | Did J. P. Morgans men add liquidity? Corporate investment, cash flow and financial structure at the turn of the Twentieth Century[END_REF] finds that firms affiliated with J. P. Morgan were less liquidity constrained, while the results of [START_REF] Becht | Does bank affiliation mitigate liquidity constraints? Evidence from Germany universal banks in the pre-World War I period[END_REF] suggest that coal-mining firms affiliated with a universal bank in Germany also faced less credit constraints. DeLong (2001) shows that the presence of directors affiliated with J.P. Morgan was associated with higher firm value. [START_REF] Van Overfelt | Do Universal Banks Create Value? Universal Bank Affiliation and Company Performance in Belgium[END_REF] find similar results for firms affiliated with universal banks in pre-World War I Belgium.

In our empirical analysis we will assume a bank relationship (i) if a bank has a direct equity stake in the firm, and/or (ii) if the firm has a bank director on its board. Firms with a bank equity stake are likely to have a close relationship with the bank. A relationship at the board level represents a continuous interaction and information exchange between the firm and the bank and may facilitate access to bank credit, even in the absence of a bank equity stake. We expect that firms with bank relationships extended more credit to their customers. We will also consider the financing role of trade credit by using firm size and firm age as proxies for access to external financing (e.g. [START_REF] Petersen | Trade credit: theories and evidence[END_REF].

Larger and older firms may grant more trade credit to customers because wellestablished firms have better access to external financing. It can also be expected that firms with more internal cash are able to extend more credit to their customers (e.g. Petersen and Rajan, 1996). We therefore expect a positive relationship between cash flow and trade credit granted.

Product quality explanations of trade credit on the other hand predict that larger and older firms will extend less trade credit to their customers than smaller and younger firms, because they have more established reputations: firms with established reputations need not use additional quality signals (e.g. [START_REF] Long | Trade credit, quality guarantees, and product marketability[END_REF][START_REF] Lee | Product risk, asymmetric information, and trade credit[END_REF].

IV. Sample and Variables

Sample

Our sample consists of Belgian firms listed on the Brussels Stock Exchange in the period 1905-1909. We focus on four major industries: coal-mining, tramways, railways and textiles. At the time, these were the four largest Belgian industries in terms of (i) the number of firms in the industry and (ii) the number of firms affiliated with a universal bank. We hand-collected financial statement data from the appendices to the Official Gazette ("Annexes au Moniteur Belge: Recueil Spécial des Actes des Sociétés"). This is the most reliable source of Belgian financial statement data for that period. All firms constituted under Belgian -9 -corporate law were legally required to publish their balance sheets and income statements in the Official Gazette [START_REF] Théate | Bank information monopoly and trade credit: do only banks have information about small business[END_REF].

Since the discretionary power of management to design the financial statement was high, there is substantial heterogeneity in the structure of the financial statements. Therefore we had to reformat the financial statements in a uniform structure. Fortunately, the law provided some guidelines about the depreciation of assets and the distribution of profits (Resteau, 1913a and1913b) and we were able to check practitioners' guides to get a better understanding of the accounting principles at the turn of the century (François, 1902 and[START_REF] François | Inventaires et Bilans dans les Sociétés par Actions[END_REF].

After removing some observations with missing or outlying values for the variables considered in the empirical analysis, our sample consists of 535 firmyear observations for 125 different firms.

Variables

Our measure of trade credit is accounts receivable scaled by total assets 1 . As for bank relationships, we consider relationships with six different banks: (i) the Société Générale; (ii) the Crédit Général Liégeois; (iii) the Banque d'Outremer;

(iv) the Banque Liégeoise; (v) the Banque de Bruxelles and (vi) the Banque Internationale de Bruxelles. These were the most important listed universal banks both in terms of total assets and the value of the industrial portfolio [START_REF] Durviaux | La Banque Mixte: Origine et Soutien de lExpansion Economique de la Belgique[END_REF]; see also [START_REF] Van Overfelt | Do Universal Banks Create Value? Universal Bank Affiliation and Company Performance in Belgium[END_REF]. The data on the bank equity stakes were collected from the "Recueil Financier", which listed the industrial portfolio of the 1 Accounts receivable are scaled by total assets instead of sales because the financial statements generally did not show sales. We do not consider accounts payable because financial statements in pre-World War I Belgium did not distinguish between trade liabilities and bank debt. For the bank directors, we collected data on the board of directors from the "Recueil Financier", a financial annual containing a variety of firm-specific information, including directorships3 . Bank directors are based on the boards of 1905. For some companies that went public after 1905, we also used the 1905 board if it was available in the Recueil Financier. If it was not available, we used the board from the year the listing started. Bank relationships were very stable anyway. [START_REF] Van Overfelt | Do Universal Banks Create Value? Universal Bank Affiliation and Company Performance in Belgium[END_REF] find for a very similar sample that most firms with a bank director in 1905 already had a bank director in 1895, and most firms without a bank director in 1905 did not have a bank director in 1895. As in Germany, the boards of Belgian firms had a dual structure, consisting of an executive board ("administrateurs") and a supervisory board ("commissaires").

While executive directors held a significant number of directorships in other firms, their supervisory peers rarely held directorships in other firms. We therefore assume that an executive director of the bank who is on the executive board of the firm is a bank director 4 . -----------------------Table 1 about here - ----------------------Table 1 shows the number of firms interlocked with each universal bank. 43 of the 125 firms in our sample were interlocked with a universal bank. Almost half of these firms were interlocked with either the Société Générale or the Crédit Général Liégois. The table also shows the number of equity stakes held by each universal bank. Most of these stakes were held by the Société Générale; the other banks had very few or no equity stakes. All the firms in which a bank had an equity stake were also interlocked with that bank.

As for the other explanatory variables, Ln(size) is the natural logarithm of total assets; and ln(age) is the natural logarithm of difference between the current year and the year the firm transformed to a limited liability company. Cash flow is operating profit plus depreciation over the fiscal year, scaled by the book value of total assets.

The financial statements of the firms considered in this study distinguish between two forms of debt: "obligations" or bonds, and "créditeurs", which is all other debt. "Créditeurs" could be bank debt but also trade liabilities. We include both Bonds and Other debt, which are scaled by the book value of total assets, as control variables in our regressions. Firms with access to bond financing and/or other debt financing may redistribute funds via trade credit to other firms. On the other hand, it is generally observed that firms try to "match maturities": shortall. Three years later, he was a director of 13 different firms. In most of these firms, Edouard Despret had been a director when he died (Sources: the 1906 annual report of the Société Générale and the 1906 and 1910 editions of the Receuil Financier). All the firms in which Edouard Despret held positions before his death still had one or two Société Générale Directors in 1909. -12 -term needs are financed with short-term debt and long-term needs are financed with long-term sources of funds (e.g. [START_REF] Stohs | The determinants of corporate debt maturity structure[END_REF][START_REF] Deloof | Trade Credit, Corporate Groups, and the Financing of Belgian Firms[END_REF]. Maturity matching suggests a positive relationship between accounts receivable and other debt, and a negative relationship between accounts receivable and bonds. To capture industry-specific trade credit policies, we also consider dummies for the coal-mining, tramways, and textiles industries.

- ----------------------Table 2 about here - ---------------------------------------------Table 3 about here -----------------------Table 2 presents descriptive statistics. Investments in accounts receivable were substantial: on average 20.1% of total assets (median is 14.0%). However, Table 3 shows that there were large differences across industries. While accounts receivable were on average more than 20% of total assets for coal-mining and textile firms, they constituted less than 10% of total assets for tramways and railways firms. Our sample reflects the importance of the coal-mining industry at the turn of the 20 th century: a majority of the firm-year observations (341) in our sample are observations for the coal-mining industry.

- ----------------------Table 4 about here -----------------------V. Regression results [START_REF] Petersen | Estimating Standard Errors in Finance Panel Data Sets: Comparing Approaches[END_REF].

In regression (1) we include a bank equity stake dummy as a measure of bank relationships. The coefficient of this variable, which is statistically significant at the 10% level, implies that the ratio of accounts receivable to total assets is 8.8% higher for firms with a bank equity stake than for firms without a bank equity stake. This result is consistent with the hypothesis that bank affiliated firms with better access to credit redistributed capital via trade credit to other firms.

Finance based models of trade credit predict that larger and older firms with better access to credit will extend more credit than smaller and younger firms. Product quality models on the other hand assume that larger and older firms extend less credit, because firms with established reputations need not use trade credit as a quality signal. In regression (1), the ln(size) coefficient is insignificant, while the ln(age) coefficient is positive and significant at the 10% level, which is consistent with the financing hypothesis, but inconsistent with the product quality hypothesis. We also find that cash flow is positively and significantly (at the 1% level) related to accounts receivable, confirming that firms with more internal cash were able to grant more credit to their customers.

Consistent with the maturity matching hypothesis, we find that accounts receivable are positively and significantly (at the 5% level) related to "other debt".

Finally, our results suggest that coal-mining firms extended more trade credit than In regression (2) we consider our second measure of bank relationships: a dummy which equals one if a firm had a bank director on its board, and zero otherwise.

However, the coefficient of this variable is insignificant.

Firms in trouble may extend trade credit to attempt to maintain their sales [START_REF] Petersen | Trade credit: theories and evidence[END_REF][START_REF] Marotta | When do trade credit discounts matter? Evidence from Italian firm-level data[END_REF]. We test this conjecture by separating cash flow into cash flow if positive and cash flow if negative: firms with a negative cash flow are more likely to be in trouble than firms with a positive cash flow. In regression (3), the coefficient of former remains positive and significant at the 1% level, while the coefficient of the latter is insignificant. The conjecture that firms in trouble extended more trade credit is therefore not confirmed.

Most of the sample firms in which a bank has an equity stake are coal-mining firms. It is therefore possible that the effect of bank equity stakes we found in regression (1) is in fact a proxy for a "coal-mining effect": maybe it was not firms with a bank equity stake but firms in the coal-mining industry which extended more trade credit. In regression (4) we re-estimated our base model for coalmining firms only (341 firm-year observations). The bank equity coefficient is again economically large (0.147), and statistically significant at the 5% level. This result suggests that even within the coal mining industry firms with a bank equity stake extended more trade credit. The findings for the other independent variables are also confirmed: ln(age), cash flow and other debt are all significant when we consider coal-mining firms only.

VI. Conclusions

This study investigates the factors correlated with accounts receivable in an environment characterized by high information asymmetries and poorly protected investors, focusing on the role of bank relationships. The results are generally consistent with the financing role of trade credit, but inconsistent with product quality explanations for trade credit. The results also confirm that firms with close bank ties redistributed capital via trade credit to other firms. However, only firms with a bank equity stake extended more trade credit. Firms with a banker on their board which did not have a bank equity stake did not provide more credit to their customers.

Most of the sample firms with a bank equity stake were coal-mining firms, in which the Société Générale held a stake. The Société Générale was the dominant universal bank in Belgium, not only before Word War I but for most of the 19 th and 20 th centuries. Historical evidence confirms that the Société Générale had a close relationship with the coal mining firms in which it held a stake. For example, [START_REF] Kurgan-Van Hentenryk | The Société Générale[END_REF] notes that in 1881 the Société Générale established the "Inspection des charbonnages" (collieries inspectorate), in order to monitor the costs of collieries under its control. De Beelde (1992) provides evidence that the Société Générale imposed cost accounting rules on affiliated coal mining firms. This raises the question which customers were granted more credit by bank affiliated firms. It cannot be ruled out that affiliated firms extended more credit only to firms controlled by the same bank. As a matter of fact, [START_REF] Deloof | Trade credit, product quality, and intragroup trade: some European evidence[END_REF] find evidence that Belgian firms belonging to a group in the 1990s used -16 -trade credit as a tool to finance other group members. The available data for the period considered in this study do not allow us to investigate trade payables policies. However, our finding that firms with a bank director, which were likely to have easier access to credit than firms without a bank director, did not extend more trade credit could be interpreted as an indication that it was not the availability of credit per se which led firms with a bank equity stake to provide more trade credit. This table reports the firm interlocks and equity stakes for each universal bank. The sample consists of 535 firm-year observations in the coal-mining, trams railways and textiles industries. A firm is interlocked with a universal bank if an executive director of the bank is also a member of the executive board of the company. All the firms in which a bank had an equity stake were also interlocked with that bank.

  point out that there are at least three sources of such cost advantages. First, the trading relationship gives suppliers an informational

  equity stakes and board positions may have provided signals of credit worthiness to capital markets (see e.g.

  this study. We are able to identify only 17 firms in which a bank had an equity stake 2 .

  errors clustered by firm account for the fact that standard errors of regression coefficients are downward biased if residuals are correlated across time for a given firm. When both fixed firm and time effects are present in the data, the time effect can be addressed by including time dummies and then estimating standard errors clustered on the other dimension

  industries: the coefficient of the coal-mining dummy is positive and significant at the 5% level.

  Table 4 reports regression results. All regressions are estimated assuming random effects. We use clustered standard errors to asses the significance of the estimated

Table 1 .

 1 Company Interlocks and Equity Stakes for each Universal Bank

Page 24 of 26 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

	Submitted Manuscript
		Number of	Number	Average
	Universal Bank	interlocked	of equity	percentage of
		firms	stakes	shares held
	Société Générale	17	12	20%
	Crédit Général Liégeois Banque d'Outremer Banque de Bruxelles Banque Liégeoise Banque Internationale de Bruxelles F o r	18 10 5 4 3	3 1 1	26% 4% 4%
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Table 2 .

 2 Descriptive StatisticsThis table reports descriptive statistics for a sample of 535 firm-year observations for 125 listed Belgian firms in the coal-mining, trams, railways and textile industries in the period 1905-1909. Accounts Receivable, Cash Flow, Bonds and Other Debt are scaled by the book value of total assets. Cash Flow is operating profit plus depreciation over the fiscal year. Size is total assets at the beginning of the fiscal year (expressed in Belgian Francs). Age is the difference between the current year and the year the firm transformed to a limited liability company.

		Mean	Median	Standard Deviation	Minimum Maximum
	Accounts Receivable	0.201	0.140	0.196	0.000	0.926
	Cash Flow	0.140	0.113	0.121	-0.255	0.871
	Size	9,077,236 4,544,958 15,932,929 291,502 135,587,143
	Age	28.250	23.000	19.328	1.000	81.000
	Bonds	0.127	0.000	0.187	0.000	0.843
	Other Debt	0.147	0.099	0.139	0.000	0.747

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Table 3

 3 Average accounts receivable / total assets for each industryThis table reports the average accounts receivable over total assets ratio for a sample of 535 firm-year observations for 125 listed Belgian firms in the coalmining, trams, railways and textile industries in the period 1905-1909. The number of observations for each industry-year are given in parentheses.

		Submitted Manuscript		Page 26 of 26
		Coal-mining	Tramways	Railways	Textile
	1905	0.233	0.097	0.081	0.233
		(66 obs.)	(13 obs.)	(6 obs.)	(12 obs.)
	1906	0.256	0.066	0.082	0.249
		(69 obs.)	(16 obs.)	(6 obs.)	(12 obs.)
	1907	0.248	0.081	0.097	0.203
		(70 obs.)	(17 obs.)	(5 obs.)	(20 obs.)
	1908	0.230	0.068	0.095	0.202
		(69 obs.)	(17 obs.)	(5 obs.)	(20 obs.)
	1909	0.226	0.061	0.092	0.196
		(67 obs.)	(18 obs.)	(5 obs.)	(22 obs.)
	1905-1909	0.239	0.074	0.089	0.210

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

Table 4 .

 4 Regression ResultsRandom effects regressions are based on a sample of 535 firm-year observations for 125 listed Belgian firms in the coal-mining, trams, railways and textile industries in the period 1905-1909. Accounts Receivable (the dependent variable), Cash Flow, Bonds and Other Debt are scaled by the book value of total assets. Cash Flow is operating profit plus depreciation over the fiscal year. Ln(size) is the natural logarithm of total assets. Ln(age) is the natural logarithm of difference between the current year and the year the firm transformed to a limited liability company. All regressions include year dummies. P-values (robust for heteroscedasticity) are in parentheses below each coefficient. P-values for random effects estimations are based on clustered standard errors. ***: denotes significance at the 1% level; **: denotes significance at the 5% level; *: denotes significance at the 10% level.

	Page
		(1)	(2)	(3)	(4)
	Sample:	All Companies All Companies All Companies	Coal-Mining Companies
	Bank Equity Stake	0.088*	-	0.088*	0.147**
		(0.076)		(0.077)	(0.019)
	Bank Director	-	0.042	-	-
			(0.240)
	Ln (Size)	-0.003	0.001	-0.003	-0.021
		(0.866)	(0.949)	(0.877)	(0.431)
	Ln (Age)	0.021*	0.019	0.021*	0.026*
		(0.059)	(0.118)	(0.059)	(0.056)
	Cash Flow	0.341***	0.336***	-	0.393***
		(0.000)	(0.000)		(0.000)
	Positive Cash Flow	-	-	0.344***	-
				(0.000)
	Negative Cash Flow	-	-	0.241	-
				(0.293)
	Bonds	0.052	0.048	0.046	0.133
		(0.610)	(0.642)	(0.687)	(0.230)
	Other Debt	0.164**	0.162**	0.162**	0.233**
		(0.011)	(0.013)	(0.014)	(0.020)
	Coal-Mining	0.152**	0.154**	0.150**	-
		(0.036)	(0.034)	(0.045)
	Tramways	0.032	0.016	0.032	-
		(0.645)	(0.811)	(0.648)
	Textiles	0.110	0.109	0.109	-
		(0.111)	(0.107)	(0.119)
	Overall R²	0.314	0.230	0.316	0.328
	No. of observations	535	535	535	341

27 of 26 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

The limited number of bank equity stakes is remarkable, as studies on the history of Belgian universal banks tend to stress the role of these banks as equity investors (see for example[START_REF] Kurgan-Van Hentenryk | Finance and Financiers in Belgium, 1880-1914[END_REF]. However, this finding is consistent with the conclusion of[START_REF] Fohlin | Finance Capitalism and Germanys Rise to Industrial Power: Corporate Finance, Governance, and Performance from the 1840s to the Present[END_REF] that German banks in the late 19th century owned few equity stakes in non-financial companies.

 3 To check the accuracy of the "Recueil Financier", we compared the information in the "Receuil Financier" with the entries in the appendices of the Official Gazette for a sub-sample of firms but we found no differences.

We cannot rule out the possibility that some of these directors were in fact industrialists sitting on the board of a bank, and not bankers on the board of industrial firms. However, anecdotal evidence suggests that these directors generally were representing the interests of the bank. For example, Edouard Despret, who died in October 1906, held board positions in 19 different firms at the time of his death. He was the vice-president of the Société Générale, which was by far the largest bank in Belgium. In November 1906, he was succeeded on the board of the Société Générale by Jean Jadot. Before becoming a director of the Société Générale, Jean Jadot held no board positions at

G3 -Corporate Finance and Governance < G -Financial Economics, G21 -Banks|Other Depository Institutions|Mortgages < G2 -Financial Institutions and Services < G -Financial Economics


Markus Leibrecht
email: markus.leibrecht@wu-wien.ac.at.

Johann Scharler
email: hann.scharler@jku.at.

Borrowing Constraints and International Risk Sharing: Evidence from Asymmetric Error-Correction

Keywords: International Risk Sharing, Error-Correction International Risk Sharing, Error-Correction JEL codes: F36, F41 International Risk Sharing, Error-Correction JEL codes: F36, F41

We analyze the adjustment process of consumption growth after disequilibrating output shocks in a sample of OECD countries. In particular, we test the hypothesis that consumption is smoothed to a lesser degree after negative shocks, whereas the impact of a positive shocks is delayed for a longer period of time. Our analysis is based on an error-correction framework that allows for asymmetric adjustment.

We find that the mean adjustment lag after a negative shock is 6.5 years compared to 10.5 years after a positive shock. This result is consistent with the interpretation that borrowing constraints limit the degree to which the impact of negative shocks on consumption can be smoothed.

In this paper we focus on this adjustment process back to the longrun relationship and test the hypothesis that the impact of negative output shocks on consumption can be smoothed to a lesser degree than the impact of positive shocks. Although the literature documents and provides explanations for the low degree of international consumption risk sharing it is rather silent concerning the adjustment process after shocks to output. Thus, we contribute to the literature in this respect and document a further role of borrowing constraints besides their relevance for the level of risk sharing.

Our analysis is based on a bivariate error-correction model (ECM) framework, which allows us not only to estimate the degree of international risk sharing in the long-run but also the speed-of-adjustment back to the longrun relationship. To study asymmetries we allow the speed-of-adjustment coefficients to depend on the sign and also the size of disequilibrating shocks.

We find that the mean adjustment lag after a negative output shock is 6.5 years compared to 10.5 years after a positive shock. Hence, the adjustment of consumption growth to output shocks is indeed asymmetric: countries can smooth the impact of adverse shocks on consumption for a shorter period than the impact of positive shocks. This result is consistent with the interpretation that borrowing constraints become binding relatively quickly after negative shocks and thereby reduce a country's ability to smooth consumption.

The paper is closely related to the literature that investigates the degree of risk sharing at different time horizons (see e.g. Becker and Hoffmann, 2006;[START_REF] Artis | Financial globalization, international business cycles and consumption risk sharing[END_REF]Canova and Ravn, 1996). These studies typically argue that the observed lack of international risk sharing may be due to a lack of insurance against permanent shocks. In contrast to our paper, this branch of the literature does not explicitly take the adjustment process into account. Methodologically, our analysis is also related to the literature that distinguishes between positive and negative changes in output to investigate 3 F o r P e e r R e v i e w the role of borrowing constraints in the context of the permanent output hypothesis (Shea, 1995a,b;[START_REF] Altonji | Testing the response of consumption to income changes with (noisy) panel data[END_REF].

The remainder of the paper is structured as follows: Section 2 outlines the empirical methodology which forms the basis of our analysis. Section 3 is devoted to data and estimation results while section 4 summarizes and concludes the paper.

Empirical Model and Methodology

As it is standard in the literature we rely on a measure of risk sharing based on the benchmark of complete markets. If markets are complete, then consumption should be (i) highly correlated across countries and (ii) not be influenced by any idiosyncratic variables, as for instance output (see Asdrubali et al., 1996;[START_REF] Sørensen | International risk sharing and European monetary unification[END_REF]. Formally: log c it = α i +log c t , where c it denotes real per capita consumption at time t in country i, c t is aggregate consumption, which is defined as a population weighted average over all relevant countries, and α i are country-specific time-invariant effects. If country-specific risks are not perfectly pooled across countries, then consumption will track idiosyncratic or country-specific income, ỹit = log y it -log y t , where y it is real per capita output in country i and y t is aggregate output. Thus, we obtain

cit = α i + β ỹit , (1)
where cit = log c 1t -log c t . Note that (1) represents a long-run relationship and therefore the dynamics of cit and ỹit can be expressed in terms of an ECM.

As noted, if countries face constraints on the adjustment of their foreign asset holdings, then the adjustment of consumption may be asymmetric after shocks to output. Specifically, running a negative net foreign asset position may only be possible in the short run and therefore consumption is likely to adjust quicker after adverse shocks. In our framework this hypothesis translates into asymmetric error-correction. We interpret periods of below average output growth as periods when adverse shocks have occurred. Similarly, periods of positive shocks correspond to above average output growth rates. To model this asymmetry, we introduce two dummy variables capturing positive, D + it , and negative, D - it , deviations from average output growth. Specifically, these dummies are defined as:

D + it = 1 if ∆ỹ it > ∆ỹ i. and D + it = 0 otherwise, where ∆ỹ i. = 1 T T i=1
∆ỹ it is the mean change in ∆ỹ it over the sample period for each country. Similarly, D - it = 1 if ∆ỹ it < ∆ỹ i. and D - it = 0 otherwise. Taking the potential asymmetry in the adjustment process into account, we estimate a bivariate ECM of the following type:

∆c it = γ 10 + γ 11 ∆ỹ it-1 + γ 12 ∆c it-1 + λ c (c it-1 -α i -β ỹit-1 ) + e 1it , (2) ∆ỹ it = γ 20 + γ 21 ∆c it-1 + γ 22 ∆ỹ it-1 + λ y (c it-1 -α i -β ỹit-1 ) + e 2it , (3)
where | in this case. Furthermore, if borrowing constraints limit the amount of consumption smoothing after negative shocks then it appears conceivable that consumption is particularly exposed to large adverse shocks. That is, constraints may be particularly tight if countries have to borrow large amounts. To capture this idea, we introduce an additional set of dummy variables which take the size of the deviation from average output growth into account: Note that our framework is targeted at the analysis of asymmetries in the adjustment process, but not of the level of risk sharing itself. This is the case as the dummy variables we introduce are defined with respect to short-run deviations from average growth. Hence, this classification cannot be used to separate countries according to the level of risk sharing they can achieve in the long run. Therefore, we restrict the β coefficient in (1) to be the same in all countries. This assumption is justified as the Hausman-test carried out by Leibrecht and Scharler (2008), using the same data and same specification, does not reject the null hypothesis of a common long-run degree of international consumption risk sharing. Intuitively, the homogeneity with respect to β is consistent with the interpretation that in the long run restrictions that may limit the amount of risk sharing have rather similar implications across countries.

λ k = λ - k D - it + λ + k D + it , for k = c,
S 0 it = 1 if ∆ỹ i. -σ < ∆ỹ it < ∆ỹ i.
To empirically implement (2) we follow the methodology outlined in Engle and Granger (1987). First, we estimate the long-run relationship in (1), and second, we estimate (2) and (3) by Seemingly Unrelated Regressions. 4

A 'general-to-specific'-approach is used to derive a parsimonious model. We thereby apply the sequential general-to-specific testing strategy proposed by 3 In principle, we can also identify periods characterized by more extreme deviations by using e.g. two standard deviations. However, since such periods rarely occur, we would not have enough observations.

4 Our approach requires the series to be cointegrated. Since evidence in favor of cointegration has already been reported in Leibrecht and Scharler (2008) for our data set, we start directly with the estimation of the ECM.

F o r P e e r R e v i e w Hall (1994) and recommended by Maddala and Kim (1998) to each single equation and each variable. We start with a lag of T 1/3 (see [START_REF] Said | Testing for unit roots in autoregressive-moving average models of unknown order[END_REF], which implies that the maximum number of lags is 4 in our case, and test down until the lag is statistically significant at the 10 percent level. 5

3 Data and Estimation Results Heston et al. (2002). We use real per capita consumption and real per capita GDP measured in constant (1996) international prices as proxies for cit and ỹit . World aggregates are calculated as weighted averages:

Data Description

y t = 21
i=1 =j w it y it and c t = 21 i=1 =j w it c it , with w it = pop it / 21 i=1 =j pop it , where pop it denotes population. Thus, c it and y it themselves are not included in the world aggregate.

Table 1 provides an overview of the deviations from average output growth in our sample. Columns 2 and 3 of the table show the number of years characterized by above and below average output growth which correspond to the definitions of D + it and D - it . The remaining columns correspond to the definitions of S 0 it , S - it and S + it . Not surprisingly, the first two columns show that the numbers of above and below average growth rates are rather balanced. Moreover, years with 5 Note, that we use the same number of observations to compare models with different lag length. The common number of observations used is 945, which is the number of observations in case of lag = 4, the maximum lag length we start with. Italy and Japan saw their last above average growth rates in 1995 and 1991, respectively, which corresponds to the weak performance these two countries experienced in the second half of the 1990s. From the last three columns we see that the vast majority of years represent 'normal' output growth, whereas periods where growth is at least one standard deviation above or below the average occur relatively rarely. Thus, the definitions of S - it and S + it capture rather extreme events. Such extreme events occurred, for instance, in the Scandinavian countries (Denmark, Sweden and Finland) which experienced severe economic downturns around 1990. Ireland and Luxembourg, in contrast, experienced relatively high growth rates in the late 1990s, which may be due to substantial gains from the European common market.

[Table 1 here ]

Results

Table 2 shows the estimated parameters of the long-run relationship (1).

We see that ỹit enters strongly significantly with a point estimate of 0.904.

As discussed in Asdrubali et al. (1996) such an estimate implies that only around 10 percent of idiosyncratic risks are pooled in the long run. Thus, we find that long-run risk sharing is rather limited, which is in line with the literature cited above.

[Table 2 What we are primarily interest in are the speed-of-adjustment coefficients in the consumption equation. From Table 3 we see that the point estimates for λ + c and λ - c are both significantly negative which shows that consumption growth reacts to deviations from the long-run relationship in times of below and also above average output growth. Comparing the magnitudes of the estimated speed-of-adjustment coefficients shows that cit reacts more strongly to deviations from the long-run relationship when output growth is below average output growth. In this case, the estimated speed-of-adjustment coefficient is -0.150. In the case that ∆ỹ it > ∆ỹ i. , we estimate the speed of adjustment coefficient to be -0.094. Furthermore, the null hypothesis of equal speed-of-adjustment coefficients for positive and negative deviations is rejected at the 10 percent level. The estimates imply mean adjustment lags of about 6.5 and 10.5 years. 6

Thus, the ordering of the estimated speed-of-adjustment coefficients is consistent with our hypothesis that borrowing constraints become binding soon after negative shocks. 7 Intuitively, when countries experience adverse 6 The mean adjustment lags are calculated as 1/ λ + c and 1/ λ - c , respectively. 7 Note that symmetric adjustment is not necessarily inconsistent with the existence of borrowing constraints. Consider a country that experiences a positive shock. To smooth the impact of the shock, this country can lend to other countries. However, if all or most other countries face credit constraints, then this may not be possible. Hence, the response to the positive shock would be similar to the response to a negative shock. Thus, symmetric adjustment also indicate that almost all countries in the sample face constraints. However, given our results, this case does not appear to be relevant empirically. shocks to their idiosyncratic output growth rates, then their consumption growth mirrors output growth closely already after a relatively short period of time. The lower panel of Table 3 shows the results for the output equation (3). We see that three of its own lags are sufficient to capture the short run dynamics of idiosyncratic output growth ∆ỹ it . The table also shows that the response of ∆ỹ it is only significantly different from zero in the case of positive deviations. Nevertheless the null that λ - y = λ + y = 0 is rejected at the 10 percent level. Hence, ∆ỹ it cannot be considered to be weakly exogenous in our system. Thus, consumption as well as output respond to shocks to restore the long-run relationship. Note that direct convergence to the long-run relationship occurs if λ c is negative and λ y is positive. We find, however, that λ + y turns out to be significantly below zero. Nevertheless, the system is stable since Juselius, 2006). The absolute value of the sum of the speed-of-adjustment coefficients is by far larger in the consumption equation.

|λ c | > |λ y | (see
[Table 3 here] Next, we characterize risk sharing not only according to the sign of the deviation from average output growth, but also according to the the size of the deviation. We re-estimate the ECM in (2) -( 3) but now we parameterize 4 shows that we obtain the lowest speed-of-adjustment coefficient in the consumption equation of -0.216 for the case where output growth is at least one standard deviation below average output growth. For values of ∆ỹ it which are close to the average ∆ỹ i. we obtain a markedly higher speed-of-adjustment coefficient of -0.111 and therefore a slower adjustment of consumption growth. The slowest adjustment is implied by the estimate for λ + c of -0.097 which is associated with output growth rates which are at least one standard deviation above average output growth. Most im- Experiencing a pronounced negative output shock results in a much faster return to the long-run equilibrium than experiencing a negative, yet 'normal', or a positive deviation from average growth. Thus, the ordering of the estimated speed-of-adjustment coefficients is again consistent with our hypothesis. Note also that our results imply that consumption growth responds rather similarly after marked positive shocks and in normal periods from both a substantive and a statistical viewpoint.

λ k = λ - k S - it + λ 0 k S 0 it + λ + k S + it , for k = c, y. Table
Turning to the results for the output equation in the lower panel of the table, we again see that ∆ỹ it reacts significantly only in the case of positive deviations. As before, the null of joint insignificance of the speedof-adjustment coefficients in ( 3) is rejected and the system is stable.

[Table 4 here]

Concluding Remarks

In this paper we analyze whether the adjustment process of consumption growth after disequilibrating output shocks depends on the sign of the shock.

We indeed find evidence in this respect. The long-run equilibrium is restored relatively quickly after adverse output shocks, whereas consumption growth adjusts slower after positive output shocks. Thus, countries are able to smooth the impact of adverse shocks only to a comparably limited extent. One explanation is that borrowing constraints on international financial markets reduce the opportunities to smooth adverse output effects. Notes: The columns labeled 'below' and 'above' report the number of periods in which real per capita GDP growth was either below or above the average growth rate. In brackets we report the last year when a positive or negative deviation occurred. The columns labeled 'positive', "negative" and "normal" report the number of years in which real per capita GDP growth was at least one standard deviation blow average growth, at least one standard deviation above average growth and within plus/minus one standard deviation of the average growth rate, respectively. In brackets, we report the last year when the respective deviation occurred. shows rather convincingly that the degree to which idiosyncratic risk is diversified internationally is low (see e.g. [START_REF] Obstfeld | The six major puzzles in international macroeconomics[END_REF]Lewis, 1999;[START_REF] Obstfeld | Are industrial-country consumption risks globally diversified?[END_REF]Backus et al., 1992).

1
The sources of this lack of international risk sharing are not yet fully understood. 2 Sørensen and Yosha (1999) find that international risk sharing among countries occurs to a large extent through international borrowing and lending. That is, after a country-specific shock, countries adjust their net asset positions to smooth the impact of the shock. However, this type of consumption smoothing may be prevented to some extent if countries face borrowing constraints on international financial markets. In this case it may not be possible to borrow sufficient amounts to smooth the impact of a negative shock. Moreover, even if countries are able to borrow, they may not be able to sustain a negative net foreign asset position for longer periods of time. In contrast, the impact of positive shocks can be delayed for a longer period of time via international lending. Thus, if countries face borrowing constraints on the adjustment of their net foreign asset position, then we expect that consumption responds stronger and also sooner to adverse shocks. This implies that constraints on international financial markets may not only lead to a low degree of risk sharing but also to an asymmetric adjustment process after disequilibrating output shocks depending on the sign of these shocks.

In this paper we focus on this adjustment process back to the long- run relationship and test the hypothesis that the impact of negative output shocks on consumption can be smoothed to a lesser degree than the impact of positive shocks. Although the literature documents and provides explanations for the low degree of international consumption risk sharing it is rather silent concerning the adjustment process after shocks to output. Thus, we contribute to the literature in this respect and document a further role of borrowing constraints besides their relevance for the level of risk sharing.

Our analysis is based on a bivariate error-correction model (ECM) framework, which allows us to estimate not only the degree of international risk sharing in the long-run but also the speed-of-adjustment back to the longrun relationship. To study asymmetries we allow the speed-of-adjustment coefficients to depend on the sign and also the size of disequilibrating shocks.

We find that the adjustment occurs significantly faster in case of negative shocks. Hence, the adjustment of consumption growth to output shocks is indeed asymmetric: countries can smooth the impact of adverse shocks on consumption for a shorter period of time than the impact of positive shocks.

This result is consistent with the interpretation that borrowing constraints become binding relatively quickly after negative shocks and thereby reduce a country's ability to smooth consumption.

The paper is closely related to the literature that investigates the degree of risk sharing at different time horizons (see e.g. Becker and Hoffmann, 2006;[START_REF] Artis | Financial globalization, international business cycles and consumption risk sharing[END_REF]Canova and Ravn, 1996). These studies typically argue that the observed lack of international risk sharing may be due to a lack of insurance against permanent shocks. In contrast to our paper, this branch of the literature does not explicitly take the adjustment process into account. Methodologically, our analysis is also related to the literature that distinguishes between positive and negative changes in output to investigate the role of borrowing constraints in the context of the permanent income hypothesis (Shea, 1995a,b;[START_REF] Altonji | Testing the response of consumption to income changes with (noisy) panel data[END_REF].

3

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

w

The remainder of the paper is structured as follows: Section 2 outlines the empirical methodology which forms the basis of our analysis. Section 3 is devoted to data and estimation results while section 4 summarizes and concludes the paper.

Empirical Model and Methodology

As it is standard in the literature we rely on a measure of risk sharing based on the benchmark of complete markets. If markets are complete, then consumption should be (i) highly correlated across countries and (ii) not be influenced by any idiosyncratic variables, as for instance output (see Asdrubali et al., 1996;[START_REF] Sørensen | International risk sharing and European monetary unification[END_REF].

Formally, assuming constant relative risk aversion utility functions, an efficient allocation is characterized by: log c it = α i + log c t , where c it denotes real per capita consumption at time t in country i, c t is aggregate consumption, which is defined as a population weighted average over all relevant countries, and α i are country-specific time-invariant effects. If country-specific risks are not perfectly pooled across countries, then consumption will track idiosyncratic or country-specific income, ỹit = log y it -log y t , where y it is real per capita output in country i and y t is aggregate output. Thus, we

obtain cit = α i + β ỹit , (1)
where cit = log c 1t -log c t . Note that (1) represents a long-run relationship and therefore the dynamics of cit and ỹit can be expressed in terms of an ECM.

As briefly mentioned above, our analysis is based on the idea that constraints on the adjustment of foreign asset holdings give rise to asymmetric adjustment processes. Specifically, running a negative net foreign asset position may only be possible over a rather short period of time. Therefore consumption is likely to adjust quicker after adverse shocks. sharing is indeed limited by binding borrowing constraints, then we expect to observe that consumption reacts faster to negative movements in output.

Note that although an asymmetric adjustment indicates that countries face borrowing constraints, a symmetric adjustment does not imply that borrowing constraints are absent or not binding. Consider the extreme case of autarky, where countries have no access to international financial markets and are therefore confined to consume their outputs in every period.

By definition, this situation corresponds to a complete lack of international risk sharing and consumption perfectly tracks income in each country. It follows that the adjustment of consumption is necessarily symmetric, regardless of the sign of any shocks that lead to movements in output. Thus, although countries face extremely severe constraints in the sense that they are completely unable to participate on international financial markets, the adjustment process is still symmetric.

Even if we consider less extreme situations, symmetric adjustment may still be consistent with the existence of borrowing constraints. Consider a country that experiences a positive shock. To smooth the impact of the shock, this country can lend to other countries. However, if all or most other countries face credit constraints, then this may not be possible. Hence, the response to the positive shock would be similar to the response to a negative shock. Thus, symmetric adjustment may also indicate that almost all countries in the sample face constraints.

In short, although an asymmetric adjustment is consistent with the interpretation that countries face borrowing constraints on international financial markets, a symmetric adjustment process does not allow for an unambiguous interpretation.

In our estimation framework, asymmetric adjustment due to constraints translates into asymmetric error-correction. We interpret periods of below average output growth as periods when adverse shocks have occurred. Sim-

∆c it = γ 10 + γ 11 ∆ỹ it-1 + γ 12 ∆c it-1 + λ c (c it-1 -α i -β ỹit-1 ) + e 1it , (2) ∆ỹ it = γ 20 + γ 21 ∆c it-1 + γ 22 ∆ỹ it-1 + λ y (c it-1 -α i -β ỹit-1 ) + e 2it , (3)
where | in this case. Furthermore, if borrowing constraints limit the amount of consumption smoothing after negative shocks then it appears conceivable that consumption is particularly exposed to large adverse shocks. That is, constraints may be particularly tight if countries have to borrow large amounts. To capture this idea, we introduce an additional set of dummy variables which take the size of the deviation from average output growth into account: S 0 it = 1 if ∆ỹ i. -σ < ∆ỹ it < ∆ỹ i. + σ and S 0 it = 0 otherwise, where ∆ỹ i. is the average change in idiosyncratic output in country i and σ its standard deviation over the sample period. Similarly, S - it = 1 if ∆ỹ it < ∆ỹ i. -σ and S - it = 0 otherwise, and S + it = 1 if ∆ỹ it > ∆ỹ i. + σ and S + it = 0 otherwise. Thus, this set of dummy variables distinguishes between times of roughly average Note that the dummy variables we introduce are defined with respect to short-run deviations from average growth. Hence, this classification cannot be used to separate countries according to the level of risk sharing they can achieve in the long run. Therefore, we restrict the β coefficient in (1) to be the same in all countries. This assumption is justified as the Hausman-test carried out by Leibrecht and Scharler (2008), using the same data and same specification, does not reject the null hypothesis of a common long-run degree of international consumption risk sharing. Intuitively, the homogeneity with respect to β implies that in the long run any restrictions that limit the amount of risk sharing have rather similar implications across countries.

λ k = λ - k D - it + λ + k D + it , for k = c,
To empirically implement (2) we follow the methodology outlined in Engle and Granger (1987). First, we estimate the long-run relationship in (1), and second, we estimate (2) and (3) by Seemingly Unrelated Regressions. 4

A 'general-to-specific'-approach is used to derive a parsimonious model. We thereby apply the sequential general-to-specific testing strategy proposed by Hall (1994) and recommended by Maddala and Kim (1998) to each single equation and each variable. We start with a lag of T 1/3 (see [START_REF] Said | Testing for unit roots in autoregressive-moving average models of unknown order[END_REF] and test down until the lag is statistically significant at the 10 percent significance level.

3 In principle, we can also identify periods characterized by more extreme deviations by using e.g. two standard deviations. However, since such periods rarely occur, we would not have enough observations for statistically identifying the effect.

4 Our approach requires the series to be cointegrated. Since evidence in favor of cointegration has already been reported in Leibrecht and Scharler (2008) for our data set, we start directly with the estimation of the ECM.

Data Description

In our analysis, we include 21 OECD countries: Australia, Austria, Belgium, Canada, Denmark, Finland, France, Greece, Ireland, Italy, Japan, Luxembourg, the Netherlands, Norway, New Zealand, Portugal, Spain, Sweden, Switzerland, the UK and the US. The remaining OECD countries are not included due to limited data availability. Data are annual and range from 1951 to 2000. All series are obtained from the Penn World Table, release 6.1 by Heston et al. (2002). We use real per capita consumption and real per capita GDP measured in constant (1996) international prices as proxies for cit and ỹit . World aggregates are calculated as weighted averages:

y t = 21
i=1 =j w it y it and c t = 21 i=1 =j w it c it , with w it = pop it / 21 i=1 =j pop it , where pop it denotes population. Thus, c it and y it themselves are not included in the world aggregate.

Table 1 provides an overview of the deviations from average output growth in our sample. Columns 2 and 3 of the table show the number of years characterized by above and below average output growth which correspond to the definitions of D + it and D - it . The remaining columns correspond to the definitions of S 0 it , S - it and S + it . Not surprisingly, the first two columns show that the numbers of above and below average growth rates are rather balanced. Moreover, years with above and below average output growth are experienced at the end of our sample period. Notable exceptions are Ireland, Finland and the United States were the last episodes of below average growth rates were experienced in earlier years, indicating the relatively favorable macroeconomic performance of these countries in the last years. Italy and Japan, in contrast, saw their last above average growth rates in 1995 and 1991, respectively, which corresponds to the weak performance these two countries experienced in the second half of the 1990s. From the last three columns we see that the vast majority of years represent periods of 'normal' output growth, whereas periods where growth is at least one standard deviation above or below the average occur relatively rarely. Thus, the definitions of S - it and S + it capture rather extreme events. Such extreme events occurred, for instance, in the Scandinavian countries (Denmark, Sweden and Finland) which experienced severe economic downturns around 1990. Ireland and Luxembourg, in contrast, experienced relatively high growth rates in the late 1990s, which may be due to substantial gains from the European common market.

[Table 1 here ]

Results

Table 2 shows the estimated parameters of the long-run relationship (1).

Estimates are based on the country fixed effects estimator (FE) as well as on the fully-modified OLS (FMOLS) estimator of [START_REF] Pedroni | Fully-modified OLS for heterogeneous cointegrated panels[END_REF] which allows us to conduct valid inference. We see that ỹit enters strongly significantly with a point estimate of about 0.900. As discussed in Asdrubali et al. (1996) such an estimate implies that only around 10 percent of idiosyncratic risks are pooled in the long run. Thus, we find that long-run risk sharing is rather limited, which is in line with the literature cited above.

[Table 2 here] Next, we estimate the ECM in (2) -(3). The testing down procedure shows that relatively parsimonious models are sufficient to capture the shortrun dynamics. The results based on our final specification are reported in Table 3. The top panel of the table shows the estimation results for the consumption equation (2). First of all, we see that ∆c it responds significantly to ∆ỹ it-1 which is inconsistent with perfect risk sharing since any influence of idiosyncratic variables -even lagged -should be completely eliminated under perfect risk sharing. 3 we see that the point estimates for λ + c and λ - c are both significantly negative which shows that consumption growth reacts to deviations from the long-run relationship in times of below and also above average output growth. Comparing the magnitudes of the estimated speed-of-adjustment coefficients shows that cit reacts more strongly to deviations from the long-run relationship when output growth is below average output growth. In this case, the estimated speedof-adjustment coefficient is -0.150. In the case that ∆ỹ it > ∆ỹ i. , we estimate the speed-of-adjustment coefficient to be -0.094. The null hypothesis that the speed-of-adjustment coefficients are equal is marginally rejected at the 10 percent significance level. The estimated coefficients of -0.094 and -0.150 translate into mean adjustment lags of 10.5 and 6.5 years in case of negative and positive income shocks. 5 Thus, these results provide first evidence in favor of our hypothesis that borrowing constraints become binding soon after negative income shocks. Intuitively, when countries experience adverse shocks to their idiosyncratic output growth rates, then their consumption growth mirrors output growth closely already after a relatively short period of time.

The lower panel of Table 3 shows the results for the output equation (3). We see that three of its own lags are sufficient to capture the short run dynamics of idiosyncratic output growth ∆ỹ it . The table also shows that the response of ∆ỹ it is only significantly different from zero in the case of positive deviations. Nevertheless the null that λ - y = λ + y = 0 is rejected at the 10 percent level. Hence, ∆ỹ it cannot be considered to be weakly exogenous in our system. Thus, consumption as well as output respond to shocks to restore the long-run relationship. Note that direct convergence to the long-run relationship occurs if λ c is negative and λ y is positive. We find, 5 The mean adjustment lags are calculated as 1/ λ + c and 1/ λ - c , respectively. however, that λ + y turns out to be significantly below zero. Nevertheless, the system is stable since |λ c | > |λ y | (see Juselius, 2006).

[Table 3 here] Next, we characterize risk sharing not only according to the sign of the deviation from average output growth, but also according to the size of the deviation. We re-estimate the ECM in ( 2) -( 3) but now we parameterize

λ k = λ - k S - it + λ 0 k S 0 it + λ + k S + it , for k = c, y.
Table 4 shows that we obtain the lowest speed-of-adjustment coefficient in the consumption equation of -0.216 for the case where output growth is at least one standard deviation below average output growth. For values of ∆ỹ it which are close to the average ∆ỹ i. we obtain a markedly higher speed-of-adjustment coefficient of -0.111 and therefore a slower adjustment of consumption growth. A similar speed-of-adjustment is implied by the estimate for λ + c of -0.097 which is associated with output growth rates that are at least one standard deviation above average output growth. Most importantly however, λ - c is significantly different from both, λ 0 c and λ + c at the 5 and 10 percent significance level. The implied mean adjustment lags are about 4, 9 and 10 years respectively. Experiencing a pronounced negative output shock results in a much faster return to the long-run equilibrium than experiencing a negative, yet 'normal', or a positive deviation from average growth. Thus, the ordering of the estimated speed-of-adjustment coefficients is again consistent with our hypothesis. Note also that our results imply that consumption growth responds rather similarly after marked positive shocks and in normal periods.

Turning to the results for the output equation in the lower panel of the table, we again see that ∆ỹ it reacts significantly only in the case of positive deviations. As before, the null of joint insignificance of the speedof-adjustment coefficients in (3) is rejected and the system is stable. [Table 4 here]

Financial Globalization and the Adjustment Process

Over the last decades, international financial markets have become increasingly liberalized and integrated. Several papers show that financial integration increases the international sharing of consumption risk (see e.g. Imbs, 2006;[START_REF] Sørensen | Home bias and international risk sharing: Twin puzzles separated at birth[END_REF]. Therefore, in this section, we study how the adjustment after shocks has changed over time along with the ongoing process of the globalization of financial markets. More specifically, we split our sample in 1980 and re-estimate the long-run relationship (1) as well as the ECM in ( 2) and (3) for the two resulting subsamples. The choice of the date at which we split the sample is motivated by [START_REF] Artis | Financial globalization, international business cycles and consumption risk sharing[END_REF] who argue that since the beginning of the 1980s, the higher integration of international financial markets has resulted in higher international risk sharing. We first test for cointegration between cit and ỹit in the two subsamples using the panel cointegration test of [START_REF] Pedroni | Critical values for cointegration tests in heterogeneous panels with multiple regressors[END_REF]. Table 5 shows that the vast majority of the tests reject the null hypothesis of no cointegration for both time periods. Thus, we may conclude that cit and ỹit are cointegrated in both subsamples.

[Table 5 here] Table 6 shows the coefficients for the long-run relationship in (1) estimated over the two subsamples. As expected, the long-run exposure of consumption growth to idiosyncratic output growth has declined. Thus, we find that long-run risk sharing improved, although the extent of risk sharing still remains low. Overall, these findings are in line with [START_REF] Artis | Financial globalization, international business cycles and consumption risk sharing[END_REF], who also document an increase in risk sharing over time.

[Table 6 here Table 7 shows the results for equation (2) with ∆c it as the dependent variable. 6 We include two lags of the consumption growth rate in both subsamples, whereas lagged values of the output growth rate are only included in the first subsample. In the later subsample, lagged values of consumption growth are sufficient to capture the short-term dynamics as indicated by our testing down strategy.

Turning to the speed-of-adjustment coefficients we see that λ + and λ - are significantly different from zero in both subsamples. Note that in the period 1950 -1979 the null of an equal speed-of-adjustment, λ + = λ -cannot be rejected. Recall, however, from the discussion in Section 2 that a symmetric adjustment process does not allow for an unambiguous interpretation in terms of the relevance of borrowing constraints. The symmetric adjustment during the first subsample may simply indicate that most countries in our sample faced severe constraints. This interpretation is also consistent with the negligible extent of long-run of risk sharing we find for the period prior to 1980.

[Table 7 here] After 1980, however, we find evidence in favor of an asymmetric adjustment. Table 7 shows that the null hypothesis of equal speed-of-adjustment coefficients is rejected at the 5 percent significance level. Hence, consumption adjusts significantly faster after negative shocks which indicates that borrowing constraints are relevant and do not allow to postpone the impact of adverse shocks for longer periods of time. Our estimates imply a mean adjustment lag of 5 years if output growth is above average, whereas the mean adjustment lag is only 3 years when output growth is below aver- age. Thus, the asymmetric speed-of-adjust is in line with the interpretation that the low degree of risk sharing may be due to borrowing constraints on international financial markets.

Concluding Remarks

In this paper we analyze whether the adjustment process of consumption growth after disequilibrating output shocks depends on the sign of the shock.

We indeed find evidence in this respect, especially since 1980. The longrun equilibrium is restored relatively quickly after adverse output shocks, whereas consumption growth adjusts slower after positive output shocks.

Thus, countries are able to smooth the impact of adverse shocks only to a comparably limited extent. One explanation is that borrowing constraints on international financial markets reduce the opportunities to smooth adverse output effects.

It has to be pointed out that borrowing constraints are perhaps not the only source of asymmetric adjustment in consumption. However, since it is well documented in the literature that risk sharing occurs primarily via the adjustment of asset holdings, frictions on international credit markets that result in borrowing constraints are likely to play a non-negligible role. Notes: The columns labeled 'below' and 'above' report the number of periods in which real per capita GDP growth was either below or above the average growth rate. In brackets we report the last year when a positive or negative deviation occurred. The columns labeled 'positive', "negative" and "normal" report the number of years in which real per capita GDP growth was at least one standard deviation blow average growth, at least one standard deviation above average growth and within plus/minus one standard deviation of the average growth rate, respectively. In brackets, we report the last year when the respective deviation occurred.

4

  Page

  y, denotes the speed-of-adjustment coefficients depending on the sign the deviation from average output growth. Symmetric adjustment occurs if λ - c and λ + c have the same signs and |λ - c | = |λ + c |. If borrowing constraints limit the degree to which consumption can be smoothed, then negative output shocks should be mirrored in consumption sooner than positive shocks and therefore we expect |λ - c | > |λ + c

  + σ and S 0 it = 0 otherwise, where ∆ỹ i. is the average change in idiosyncratic output in country i and σ its standard deviation 5 period. Similarly, S - it = 1 if ∆ỹ it < ∆ỹ i. -σ and S - it = 0 otherwise, and S + it = 1 if ∆ỹ it > ∆ỹ i. + σ and S + it = 0 otherwise. Thus, this set of dummy variables distinguishes between times of roughly average or normal output growth and episodes of relatively low and high growth, where we consider values of ∆ỹ it which fall within one standard deviation from the average change as normal.3

  average output growth are experienced at the end of our sample period. Notable exceptions are on the one hand Ireland, Finland and the United States were the last episodes of below average growth rates were experienced in earlier years, indicating the relatively favorable macroeconomic performance of these countries in the last years. On the other hand

  λ - c is significantly different from both, λ + c and λ 0 c . The implied mean adjustment lags are about 4.5, 9 and 10 years respectively.

  It has to be pointed out that borrowing constraints are perhaps not the only source of asymmetric adjustment in consumption. However, since it is well documented in the literature that risk sharing occurs primarily via the 11 holdings, frictions on international credit markets that result in borrowing constraints are likely to play a non-negligible role. ., 1994. Are industrial-country consumption risks globally diversified? In: Leiderman, L., Razin, A. (Eds.), Capital mobility: the impact on consumption, investment and growth. Cambridge University Press, Cambridge. Obstfeld, M., Rogoff, K., 2000. The six major puzzles in international macroeconomics. In: Bernanke, B. S., Rogoff, K. (Eds.), NBER Macroeconomic Annual 2000. MIT Press, Cambridge, Massachusetts. Said, E., Dickey, D. A., 1984. Testing for unit roots in autoregressive-moving average models of unknown order. Biometrika 71 (3), 599-607. Scorcu, A. E., 1998. Consumption risk-sharing in Italy. Applied Economics 30 (1), 407-414. Shea, J., 1995a. Myopia, liquidity constraints, and aggregate consumption: A simple test. Journal of Money, Credit and Banking 27 (3), 798-805. Shea, J., 1995b. Union contracts and the life-cycle/permanent-income hypothesis. American Economic Review 85 (1), 186-200. Sørensen, B. E., Yosha, O., 1999. International risk sharing and European monetary unification. Journal of International Economics 45, 211-238. Stockman, A. C., Tesar, L., 1995. Tastes and technology in a two-country model of the business cycle: explaining international comovements. American Economic Review 85, 168-185

  diversified portfolios, agents should be able to pool country-specific risks and thereby reduce the impact of output fluctuations to some extent. The empirical literature on international risk sharing

  y, denotes the speed-of-adjustment coefficients depending on the sign the deviation from average output growth. Symmetric adjustment occurs if λ - c and λ + c have the same signs and |λ - c | = |λ + c |. If borrowing constraints limit the degree to which consumption can be smoothed, then negative output shocks should be mirrored in consumption sooner than positive shocks and therefore we expect |λ - c | > |λ + c

  growth and episodes of relatively low and high growth, where we consider values of ∆ỹ it which fall within one standard deviation around the average change as normal.3

  What we are primarily interest in are the speed-of-adjustment coefficients in the consumption equation. From Table

  11

Table 1 :

 1 Deviations from average output growth

	below	above	positive	negative normal
	Australia 22 (2000) 27 (1999)	3 (1958)	9 (1990)	37
	Austria 28 (2000) 21 (1998)	7 (1974)	4 (1984)	38
	Belgium 24 (1998) 25 (2000)	6 (1991)	8 (1993)	35
	Canada 19 (1996) 30 (2000)	4 (1999) 11 (1991)	34
	Switzerland 28 (1999) 21 (2000)	8 (1989)	4 (1976)	37
	Denmark 29 (1999) 20 (2000) 11 (1994)	5 (1989)	33
	Spain 23 (2000) 26 (1999)	4 (1995)	6 (1996)	39
	Finland 23 (1993) 26 (2000)	7 (1982)	7 (1992)	35
	France 28 (2000) 21 (1998)	9 (1982)	4 (1993)	36
	United Kingdom 26 (1999) 23 (2000)	8 (1994)	5 (1981)	36
	Greece 27 (1999) 22 (2000) 12 (1975)	7 (1987)	30
	Ireland 23 (1986) 26 (2000)	8 (2000) 10 (1986)	31
	Italy 24 (2000) 25 (1995)	8 (1980) 10 (2000)	31
	Japan 25 (2000) 24 (1991)	9 (1970)	6 (1999)	34
	Luxembourg 26 (1996) 23 (2000)	9 (1997)	5 (1975)	35
	Netherlands 24 (2000) 25 (1999)	5 (1991)	6 (1981)	38
	Norway 23 (2000) 26 (1997)	4 (1993) 10 (1999)	35
	New Zealand 23 (1998) 26 (2000)	6 (1991)	5 (1977)	38
	Portugal 24 (2000) 25 (1999)	6 (1991)	7 (1994)	36
	Sweden 24 (1997) 25 (2000)	7 (1982)	8 (1993)	34
	United States 23 (1991) 26 (2000)	2 (1984) 10 (1981)	37

Table 2

 2

		: Long-run Relationship
		coef	sd	p-value
	const	0.118 0.012	0.000
	ỹit	0.904 0.015	0.000
	adj. R 2 0.949
	F-Test			0.000
	obs	1050		0.000
		15

Notes: Country dummies are jointly significant and therefore included in the estimation.

Table 3

 3 Notes: The top panel of the table reports estimated coefficient for the error correction model with ∆cit as the dependent variable. The bottom panel reports the results for ∆ỹit as the dependent variable. Country dummies are jointly significant and therefore included in the estimation.Notes: The top panel of the table reports estimated coefficient for the error correction model with ∆cit as the dependent variable. The bottom panel reports the results for ∆ỹit as the dependent variable. Country dummies are jointly significant and therefore included in the estimation.

	: Asymmetric Error-Correction
	Dep. Var.: ∆c it	coef	sd	p-value
	const	-0.005 0.005	0.327
	∆ỹ it-1	0.104 0.041	0.011
	λ + c λ -c	-0.094 0.023 -0.150 0.023	0.000 0.000
	obs	966
	H 0 : λ + c = λ -c			0.089
	Dep. Var.: ∆ỹ it	coef	sd	p-value
	const	-0.005 0.004	0.168
	∆ỹ it-1	0.190 0.030	0.000
	∆ỹ it-2	-0.012 0.027	0.643
	∆ỹ it-3	0.058 0.026	0.024
	λ + y	-0.036 0.017	0.035
	λ -y	-0.009 0.017	0.577
	obs	966
	H 0 : λ + y = λ -y = 0			0.091

16 Page 16 of 38 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

5 Page 22 of 38 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

  it is the mean change in ∆ỹ it over the sample period for each country. Similarly, D - it = 1 if ∆ỹ it < ∆ỹ i. and D - it = 0 otherwise. Taking the potential asymmetry in the adjustment process into account, we estimate a bivariate ECM of the following type:

	Submitted Manuscript
	ilarly, periods of positive shocks correspond to above average output growth
	rates. To model this asymmetry, we introduce two dummy variables captur-
	ing positive, D + it , and negative, D -it , deviations from average output growth.
	Specifically, these dummies are defined as: D + it = 1 if ∆ỹ it > ∆ỹ i. and
	D + it = 0 otherwise, where ∆ỹ i. = 1 T	T i=1 ∆ỹ
	F o r
	P
	e
	e r
		R
		e
		v i e
		w

Table 1 :

 1 Deviations from average output growth

	below	above	positive	negative normal
	Australia 22 (2000) 27 (1999)	3 (1958)	9 (1990)	37
	Austria 28 (2000) 21 (1998)	7 (1974)	4 (1984)	38
	Belgium 24 (1998) 25 (2000)	6 (1991)	8 (1993)	35
	Canada 19 (1996) 30 (2000)	4 (1999) 11 (1991)	34
	Switzerland 28 (1999) 21 (2000)	8 (1989)	4 (1976)	37
	Denmark 29 (1999) 20 (2000) 11 (1994)	5 (1989)	33
	Spain 23 (2000) 26 (1999)	4 (1995)	6 (1996)	39
	Finland 23 (1993) 26 (2000)	7 (1982)	7 (1992)	35
	France 28 (2000) 21 (1998)	9 (1982)	4 (1993)	36
	United Kingdom 26 (1999) 23 (2000)	8 (1994)	5 (1981)	36
	Greece 27 (1999) 22 (2000) 12 (1975)	7 (1987)	30
	Ireland 23 (1986) 26 (2000)	8 (2000) 10 (1986)	31
	Italy 24 (2000) 25 (1995)	8 (1980) 10 (2000)	31
	Japan 25 (2000) 24 (1991)	9 (1970)	6 (1999)	34
	Luxembourg 26 (1996) 23 (2000)	9 (1997)	5 (1975)	35
	Netherlands 24 (2000) 25 (1999)	5 (1991)	6 (1981)	38
	Norway 23 (2000) 26 (1997)	4 (1993) 10 (1999)	35
	New Zealand 23 (1998) 26 (2000)	6 (1991)	5 (1977)	38
	Portugal 24 (2000) 25 (1999)	6 (1991)	7 (1994)	36
	Sweden 24 (1997) 25 (2000)	7 (1982)	8 (1993)	34
	United States 23 (1991) 26 (2000)	2 (1984) 10 (1981)	37

Table 2

 2 Notes: The table shows the estimated coefficient of ỹit in the long-run relationship, based on the country fixed effects estimator (FE) and the fully-modified OLS (FMOLS) estimator. Common time effects are subtracted (results with common time effects not subtracted are qualitatively similar). Based on one-sided, unweighed tests with common time effects subtracted (tests with common time effects not subtracted show similar results). All test statistics follow asymptotically standard normal distribution under the null of unit root or no cointegration * * * ( * * )[ * ] stands for 1% (5%) [10%].Notes: The table shows the estimated coefficient of ỹit in the long-run relationship, based on the country fixed effects estimator (FE) and the fully-modified OLS (FMOLS) estimator. Common time effects are subtracted (results with common time effects not subtracted are qualitatively similar). The table shows estimated coefficients for the error correction model with ∆c it as the dependent variable. Country dummies are significant and thus included for the 1950 -1979 period.

		: Long-run Relationship
	Dep. Var.: cit coef	sd	p-value
	FE	0.904 n.a.	n.a.
	FMOLS	0.900 0.02	0.000
	obs	1050
		18

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

21 Page 38 of 38 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

A related branch of the literature studies risk sharing within countries (e.gAsdrubali et al., 1996;[START_REF] Scorcu | Consumption risk-sharing in Italy[END_REF] Buettner,

2002;Borge and Matsen, 2004). 2 Several Explanations have been proposed in the literature. See e.g.Backus et al. (1992),[START_REF] Obstfeld | Are industrial-country consumption risks globally diversified?[END_REF],Lewis (1996),[START_REF] Stockman | Tastes and technology in a two-country model of the business cycle: explaining international comovements[END_REF],Baxter and Crucini (1995),Imbs (2006) and Hoffmann (2008).

Note that we only report the estimation results for the case where the speed-ofadjustment coefficient depends on the sign of the deviation of idiosyncratic output growth from its average but not on the size of the deviation. Taking the size of the deviation into account leaves our conclusions qualitatively unaffected. Also note that the system is stable in both time periods. However, to preserve space we do not report the results for (3) estimated over the subsamples. Detailed estimation results are available upon request.