working PAPEr n° 75

An EArly-wArning And dynAmic ForEcAsting FrAmEwork oF dEFAult ProbAbilitiEs

For thE mAcroPrudEntiAl Policy indicAtors ArsEnAl

working PAPEr n° 75

JUNE 2012

2, boulevard Royal L-2983 Luxembourg

Tél.: +352 4774-1 Fax: +352 4774 4910 www.bcl.lu • info@bcl.lu

Xisong JiN Francisco NadaL dE SimoNE

for the Macroprudential Policy Indicators Arsenal

Xisong Jin Francisco Nadal De Simone

Luxembourg School of Finance Banque centrale du Luxembourg

Draft: June 13, 2012

Abstract

The estimation of banks’ marginal probabilities of default using structural credit risk models can be enriched incorporating macro-financial variables readily available to economic agents. By combining Delianedis and Geske’s model with a Generalized Dynamic Factor Model into a dynamic t-copula as a mechanism for obtaining banks’

dependence, this paper develops a framework that generates an early warning indicator and robust out-of-sample forecasts of banks’ probabilities of default. The database comprises both a set of Luxembourg banks and the European banking groups to which they belong. The main results of this study are, first, that the common component of the forward probability of banks’ defaulting on their long-term debt, conditional on not defaulting on their short-term debt, contains a significant early warning feature of interest for an operational macroprudential framework driven by economic activity, credit and interbank activity. Second, incorporating the common and the idiosyncratic components of macro-financial variables improves the analytical features and the out-of-sample forecasting performance of the framework proposed.

JEL Classification: C30, E44, G1

Keywords: financial stability; macroprudential policy; credit risk; early warning indicators;

default probability, Generalized Dynamic Factor Model; dynamic copulas; GARCH.

The views expressed in this paper are the authors’ and do not necessarily represent those of the BCL or its staff. We thank participants at the BCL Financial Stability Review seminar in April 2012 for their comments, in particular Paolo Guarda, Abdelaziz Rouabah and Dominique Marchal as well as the FNR for its financial support. All errors and omissions remain our own. Correspondence can be sent to: Xisong Jin, Luxembourg School of Finance, 4 Rue Albert Borschette L-1246 Luxembourg, Tel: (+352) 46 66 44 5626; E-mail:

xisong.jin@uni.lu; Francisco Nadal De Simone, Banque centrale du Luxembourg, 2, boulevard royal L-2983 Luxembourg, Tel: (+352) 4774-4518; E-mail: Francisco.NadaldeSimone@bcl.lu.

I. Motivation

A relatively broad characterization of the objective of macroprudential policy is to limit systemic risk so as to minimize the costs of financial instability on the economy (ECB, June 2010). The literature on financial system risk has made a distinction between three different sources of systemic risk (ECB, December 2009): first, the exposure of all financial institutions to common, simultaneous macro-financial shocks; second, the sequential contagion from an idiosyncratic shock affecting a financial institution that spreads to other financial institutions and eventually to the real sector of the economy and; third, financial imbalances that build up over time and may unravel in a disorderly manner. Limiting financial systemic risk requires having indicators that provide a measure, albeit “fuzzy”, of financial stability, and a set of instruments to maintain and restore financial stability, when it is perturbed (Borio and Drehmann, 2009). Like the sources of systemic risk, indicators of systemic risk cover the cross-sectional dimension of systemic risk (e.g., Segoviano and Goodhart, 2009) and the time-dimension of systemic risk (e.g., Borio and Lowe, 2002). This paper contributes to several strands of the literature on both dimensions of systemic risk. Its objective is to develop a framework that generates an early warning indicator of overall credit risk in the banking sector that identifies as early as possible the build up of endogenous imbalances; that recognizes exogenous shocks timely; that factors in some manner dependence among financial institutions and; that provides robust out-of-sample forecasts of probabilities of default.¹ One of the biggest challenges for credit risk models is modelling dependence between default events and between credit quality changes. Dependence modelling is necessary to understand the risk of simultaneous defaults, the ensuing distribution of losses and the effects on financial stability. Failing to account for dependence, therefore, underestimates potential losses (Lando, 2004). This is crucial for meaningful stress testing exercises, for instance, as well as more generally, for the development of measures of systemic risk. To incorporate dependence, there are basically three broad approaches or mixtures of them: (1) to let probabilities of default be affected by common underlying observable variables; (2) to let probabilities of default be affected by underlying latent variables and; (3) to let direct contagion from a default event affect other firms. However, whether by using a mixture of distributions to model dependence or by using copula or network analysis, models require the estimation of default probabilities as a first step. This study uses two of the structural credit risk models studied in Jin and Nadal De Simone (2011a) and Jin et al (2011b), Merton (1974) model and Delianedis and Geske (2003) model, to estimate implied neutral probabilities of

1 The issue of financial institutions’ contributions to systemic risk is not addressed in this paper, but in an accompanying study.

default. To model dependence among financial institutions’ default probabilities, this paper uses the Generalized Dynamic Factor Model (GDFM) of Forni et al (2005), which has been used extensively to exploit the information from a large dataset and also for forecasting (e.g., Kabundi and Nadal De Simone, 2011, De Nicolò and Lucchetta (2012), and D’Agostino and Giannone, forthcoming).² However, Forni et al (2003) forecasting method is not easily applicable to a large number of underlying assets simultaneously, and does not generate the distributions of forecasts. As a result, this paper introduces a novel approach that combines the GDFM with a dynamic t-copula to improve the GDFM forecasting capacity.

Copula theory provides an easy way to deal with (otherwise) complex multivariate modeling (Jin and Lehnert, 2011). The advantage of the copula approach is its flexibility, because the dependence structure can be separated from the univariate marginal components, and hence the dependence structure between these marginal components can be modeled in the second stage, after the univariate distributions have been calibrated. Therefore, the copula approach provides a robust and consistent method to estimate banks’ dependence. Correlation analysis, which usually refers to linear correlation, depends on both the marginal distributions and the copula, and is not a robust measure given that a single observation can have an arbitrarily high influence.

The conditional dynamic t-copula is relatively easy to construct and simulate from multivariate distributions built on marginals and dependence structure. In fine, the GARCH-like dynamics in the copula variance and rank correlation offers multi-step- ahead predictions of the estimated GDFM common and idiosyncratic components simultaneously.

The framework of this study, therefore, shares the main core features suggested for an appropriate measure of systemic risk according to Schwaab et al (2010): a broad definition of systemic risk such as the ECB’s, an international focus, the incorporation of macroeconomic and financial conditions, unobserved factors, and the calculation of probabilities of defaults.

The main results and contributions of this paper to the time-dimension of systemic risk are, first, to show that the common component of the forward probability of banks’

defaulting on their long-term debt, conditional on not defaulting on their short-term debt, contains a significant early warning feature of interest for an operational macroprudential

2 Mechanisms for obtaining default dependence are versions of, and possible mixtures of three issues: (1) PDs are influenced by common observable variables and there must be a way of linking the joint movement of a reduced set of factors and how PDs depend on them; (2) PDs depend on unobserved background variables and credit events result in an update of the latent variables which updates PDs and; (3) direct contagion from a credit event.

framework mostly driven by measures of economic activity, credit growth and interbank activity. This is in the tradition recently surveyed by Frankel and Saravelos (2010). As such, the proposed framework measures the relative riskiness of the system in a non- crisis mode. It becomes a useful macroprudential gauge to help policymakers decising when and what remedial actions to take as systemic risk increases over time. Second, that incorporating the common and the idiosyncratic components of macro-financial variables improves the analytical features of the framework proposed, in agreement with recent work by Koopman et al (2010) and Schwaab et al (2010). Finally, and a novel contribution, the paper’s framework produces robust out-of-sample forecasting of overall banking sector credit risk, especially at the individual bank level.

The remainder of the study is organized as follows. Next section introduces a novel integrated modeling framework, and explains how to combine the GDFM with a dynamic t-copula into a dynamic forecasting framework of default probabilities. Section III discusses the data, and section IV examines the empirical results. Section V concludes.

II. An Integrated Modeling Framework

The purpose of this framework is to break down complex information into several smaller, less complex and more manageable sub-tasks that are solvable by using existing tools, and then combining their solutions in order to solve the original problem.

The decomposition approach is frequently used in statistics, operations research and engineering. For example, decomposition of time series is considered to be a practical way to improve forecasting. The usual decomposition into trend, cycle, seasonal and irregular components was motivated mainly by business analysts, who wanted to get a clearer picture of the state of the economy (Fisher, 1995). Ideally, the selected models are expected to be integrated into the same theoretical framework. However, the models are developedto solve specific questions from different strands of literature. For instance, the structural credit risk model is a model for assessing credit risk typically developed from option pricing literature; dynamic factor models have become a standard econometric tool to perform factor analysis on large datasets for measuring comovements in, and forecasting of macroeconomic time series; copulas are a fundamental tool for modeling multivariate distributions, which is extremely useful for active risk management. Despite the difficulties involved in integrating these models, there are already some examples of using the existing tools together one way or another.

For example, De Nicolò and Lucchetta (2012) use a dynamic factor model with many predictors combined with quantile regression techniques. Alessi, Barigozzi and Capasso (2007a&b) propose two new methods for volatility forecasting, which combine the GDFM

and the GARCH model, and have been proved to outperform the standard univariate GARCH in most cases by exploiting cross-sectional information.

This paper presents an integrated framework which examines credit risk emanating from the macro environment and from banks’ interconnectedness. The first contribution of the paper is to generate an early warning framework that in the tradition of Borio and Lowe (2002), associates the buildup of banking sector vulnerabilities with the real economy cycle and credit growth. The text graph illustrates the information flow of the first part of this framework:

1. The market-based / book value-based Merton and Delianedis and Geske models are used to filter out the tail risks or the probabilities of distress (probabilities of default or PDs, and distance to default or DD);

2. The probabilities of distress together with a large database of macro-financial variables are decomposed into common components and idiosyncratic components by the GDFM;

3. Those components are then broken down into their means and volatilities by the marginal dynamics of AR-GARCH models. For the in-sample estimation, a zero mean is assumed for the common components in order to keep their multi-step- ahead prediction from the GDFM;

4. The standardized residuals from the marginal dynamics, which are iid(0,1) usually with skewness and fat tails, are glued together by a dynamic t-copula with a multivariate GARCH structure;³

3 The converse of Sklar’s theorem implies that it is possible to couple together any marginal distributions, of any family, with any copula function and a valid joint density will be defined. The corollary of Sklar’s theorem is that it is possible to extract the implied copula and marginal distributions from any joint distribution (Nelsen, 1999). This framework alleviates the statistical inefficiency associated with the fact that PDs (and DD) are generated regressors.

5. By the copula approach, the standardized residuals can be further decomposed into two subsets of information: (i) information of each random variable; i.e., the marginal distribution of each variable; and (ii) information about the dependence structure (nonlinear) among the random variables.

The second main contribution of this paper is to advance a dynamic forecasting framework of credit risk for each bank by simulation from multivariate distributions built on marginal distributions and dependence structure. As shown by the following text graph, the simulation is actually an information loading process through the dynamic structures built in the first step. The forward dependence information is first generated from a multi-student’s t copula, and then marginal information is loaded up to get the forward standardized residuals. The forecasted common components and idiosyncratic components are projected by plugging-in marginal dynamics which enables customizing the information of means and volatility clusters. Last, the forecasted marginal credit risk measures are the sum of these two components. Thus, reverse engineering uncovers the tail risk or the probabilities of distress by using not only information from individual banks, but also from a large data set of macro-financial variables.

The remainder of this section reviews the methodological and statistical approaches used to estimate credit risk. First the selected models to estimate default probabilities are briefly described, and then the GDFM to nest macro-financial variables is outlined.

Last, the multivariate GARCH techniques are extended into the t-copula to introduce the dynamic forecasting framework.

1. Selected models to estimate default probabilities

In order to develop tools to measure and assess financial stability it is necessary to characterize instability. Approaches to deal with instability include, for instance,

modelling financial institutions’ default, analysing the financial system using extreme value theory, and allowing for episodes of market illiquidity. The approach taken in this study instead is to apply contingent claim analysis to the analysis and measurement of credit risk, or as it is commonly referred to, structural credit risk modeling. Structural credit risk models attempt to assess the creditworthiness of a firm by modeling the evolution of the firm’s asset values as a stochastic process, and by viewing bankruptcy as an endogenous random event linked to the value of the firm’s assets.

1.1. The Merton Model

In the Merton model, equity owners are viewed as holding a call option on the firm’s value after outstanding liabilities have been paid off. They have the option to default if the firm’s asset value falls below the present value of the notional amount—or book value—of outstanding debt (“strike price”) owed to bondholders at maturity. In other words, when the market value of the firm’s assets is less than the strike price, the value of equity is zero. Similarly, bond holders are viewed as writing a European put option to equity owners, who hold a residual claim on the firm’s asset value if the firm does not default. Bond holders receive a put option premium in the form of a credit spread above the risk-free rate in return for holding risky corporate debt (and bearing the potential loss) due to equity owners’ limited liability.

According to the Merton model, the market value of a firm’s underlying assets follows a geometric Brownian motion (GBM) of the form:

dW V dt V

dV_A=

µ

_A +

σ

_{A A}

where V_A is the firm’s assets value, with an instantaneous drift µ (the expected rate of return of the firm), and an instantaneous asset return volatility

σ

_A. W is a standard Wiener process. If X is the book value of the debt which is due at time T, Black and Scholes’ formula provides the market value of equity, V_E^:

) ( )

( d

₁

Xe

^{( )}

N d

₂

N

V

V

_E

=

_A

−

^−rT−t

where

t T

t T X r

V d

A A A

−

− +

+

= σ

σ )( ) 2

( 1 )

ln( ²

1 , d2 =d1−

σ

_A T−t , r is the risk-free rate, and N() is the cumulative density function of the standard normal distribution. The PD of the firm is the probability that its assets’ value will be less than the book value of its liabilities. The

corresponding implied neutral PD is

π

_N

= N ( − d

₂

)

. The “actual” PD is π_A=N(−DD)^, where the distance-to-default, DD, is simply the number of standard deviations that the firm is away from default:

t T

t X T

V DD

A A A

−

−

− +

=

σ

σ

µ

)( )

2 ( 1 )

ln( ²

To get to “actual” PDs from neutral PDs, the latter must be adjusted by the market price of risk, which is estimated using a capital-asset pricing model in this study:

r r u

u _M

M M A

A − +

= _, ( )

σ

ρ σ ^{, where}

σ

M is market asset return volatility, and

ρ

_A,M ^{is the}

correlation between firm’s asset return and market asset return. Alternatively, historical recovery rates can be used to move from risk neutral to “actual” PDs. The derived

“actual” PDs, however, could be still much higher than the observed PDs, the so-called

"credit spread puzzle" (Huang and Huang, 2003). Moody’s KMV, instead, maps distance to default (DD) into historical default probabilities. Chen, Collin-Dufresne and Goldstein (2009) try to adjust the level by calibrating the pricing kernel to equity returns and aggregate consumption. Fortunately, the rankings are more meaningful than the levels, given the objective of this study. A complication of CCA to calculate PDs is that the dynamics of the underlying asset value is not directly observable. To calculate

σ

_A^,

Moody’s KMV iterative procedure is used.⁴ For quoted financial institutions, the KMV approach implies taking daily equity data from the past 12 months to calculate historical assets volatility.⁵ Regarding the value of debt, the KMV approach takes all debt due in one year, plus half of the long-term debt. The KMV method is a simple two-step iterative algorithm to solve for assets volatility. The procedure uses an initial guess for volatility to determine the asset value and to de-lever the equity returns. The volatility of the resulting asset returns is used as the input to the next iteration of the procedure which, in turn, determines a new set of asset values and hence a new series of asset returns. The procedure continues in this manner until it converges.

1.2. The Delianedis and Geske Compound Option-based Risk Model

Debt maturity influences liquidity risk and PDs. However, Merton model and most credit risk models consider only a single debt maturity. This is an important drawback for a

4 Duan et al, (2004) show that the KMV estimates are identical to maximum likelihood estimates (MLE).

5 See next sub-section for the approach followed in the case of Luxembourg banks for which quoted stock prices or options on stock are not available.

central bank or a supervisor interested in assessing and tracking banks’ solvency. Geske (1977) and Delianedis and Geske (2003) consider a multi-period debt payment framework to which they apply compound option theory. This enables to account for the influence of the time structure of debt on the estimated PD.

Assume that a bank has long term debt,M₂, which matures at date T₂, and short term debt, M₁, which matures at date T₁. Between T₁andT₂, the Merton model is valid as the bank’s equity equals a call option giving the shareholder the right to buy the bank at the second payment date, T₂, by paying the strike price M₂. If at dateT₁, the call option with the bank’s value V equals at least the face value of the short term debt,M₁^:

) ( )

( _{2 2 1 2} ^{( )} ₂

1 VN k T T M e ^{1 2 1} N k

M = +

σ

_A − − ^{−rF T−T}

then the bank can roll over its debt. So, the refinancing problem, the right to buy the simple call option of the second period by paying the strike price at the first payment date, is exactly a compound option as follows:

) ( )

; , ( )

; ,

( _{1 1 2 2 2} ^{( )} _{2 1 2 1} ^{( )} ₁

2

1 1

2 2 N k k M e N k

e M t

T k

t T k

N V

V_E = _A +

σ

_A − +

σ

_A −

ρ

− ^{−rF T −t}

ρ

− ^{−rF T−t} where

t T

t T

−

= −

1

ρ

2 ^, N₂()is a bivariate cumulative normal distribution, and,

t T

t T V r

V k

A

A A F

−

−

− +

=

1

1 2

1

) 2 )(

( 1 )

ln( 1

σ

σ

, T t

t T M r

V k

A

A F

A

−

−

− +

=

2

2 2 2

2

) 2 )(

( 1 )

ln( ₂

σ

σ

.

The richness of the model allows to calculate the following risk neutral PDs: (1) the total or joint probability of defaulting at either date T₁ or date T₂, i.e., 1−^N₂(^k₁,^k₂;ρ)^{; (2) the} short-run probability of only defaulting on the short-term debt at date T₁ , i.e.,

) (

1−N k₁ and; (3) the forward probability held today of defaulting on the long-term debt at date T₂ , conditional on not defaulting on the short-term debt at date T₁ , i.e.,

) (

)

; , 1 (

1 2 1 2

k N

k k

N ρ

− . Similar to the Moody’s KMV iterative procedure, the Delianedis and Geske model is estimated by the two-step iterative algorithm. Regarding the maturity of the debt value, this study takes all short term obligations due in one year as a one-year maturity debt, and all long-term debt as a ten-year maturity debt.

1.3. The Book Value-Based Merton and Delianedis and Geske Models

As Luxembourg bank subsidiaries are not publicly quoted, an alternative approach to calculate PDs has to be followed. Hillegeist at al. (2004) demonstrate that the market- based Merton’s PD provides significantly more information about the probability of bankruptcy than do the popular accounting-based measures. However, Bharath and Shumway (2008) also examine the accuracy and PDs forecasting performance of the Merton model and find that most of its predictive power comes from its functional form rather than from the estimation method: the firm’s asset value, its asset risk, and its leverage. In an application to Brazilian and Mexican banks, Souto et al (2009) and Blavy and Souto (2009), respectively, show that the book-based Merton’s credit risk measures are highly correlated with market-based Merton’s credit risk measures.⁶ This suggests that banks’ financial statements are a crucial piece of information when forming market expectations about the probability of banks’ default. Regarding the estimation of volatility, in empirical work, a dynamic volatility model is often preferred in order to track risks more timely. However, most dynamic volatility models require many more data points than are available for Luxembourg banks. The RiskMetrics (RM) filter/model instead assumes a very tight parametric specification. The book value asset RM variance can be defined as:

Bt B

t B t t

B V V h

h + = −

ζ

₋ 2+

ζ

1 (1 )(ln( / 1))

where the variance forecast h_t^B₊₁ for period t+1 is constructed at the end of period t using the square of the return observed at the end of period t as well as the variance on period t. Although the smoothing parameter ζ may be calibrated to best fit the specific historical returns, RiskMetrics often simply fixes it at 0.94. To avoid the calibration difficulties for our limited data points, ζ is also assumed to be same for all banks, and estimated by numerically optimizing the composite likelihoods (Varin et al, 2011), here the sum of quasi maximum likelihood functions of the estimation sample over all banks simultaneously:

) / ) / ( ) 2 (ln(

) 1

( _{, 1,} ² _,

1 , 1

i t B

i t B i t T

t

i t N

i

h V

V h

QMLE ₋

=

=

+

−

=

∑ ∑

ζ

6 See also Gray and Jones, 2006, for an early application of this idea.

where N is number of banks, and there is a time series of T observations for each banks.

The recursion is initialized by setting the initial

σ

₀^B equal to the first year book value asset volatility, and the means of quarterly assets returns in a large sample are assumed to be zeros to avoid the noises brought by the sample means to the RM variance process. The estimated value of ζ _{is 0.83.}

In order to have a more forward-looking measure, the variance forecast Bt+1

σ

^{can be}

used to calibrate PDs at time t. The book-value risk neutral PDs of the Merton model can be directly estimated by:

).

) 2 )(

( 1 ) / ln(

(

2

t T

t T r

X V N

B

B B

B −

−

− +

−

=

σ

σ π

Similarly the three book-value risk neutral PDs of the Delianedis and Geske model can be estimated by substituting V_B and σ_{B into} k1and k₂ in the Geske model. Given σ_B^, the critical book value of total assets V^Bat T₁ is calculated first. Similarly, this study takes all short term debt due in one year as a one-year maturity debt, and all long-term debt as a ten-year maturity debt.

2. The Generalized Dynamic Factor Model

In recent years, large-dimensional dynamic factor models have become popular in empirical macroeconomics. The GDFM enables the efficient estimation of the common and idiosyncratic components of very large data sets. The GDFM assumes that each time series in a large data set is composed of two sets of unobserved components. First, the common components, which are driven by a small number of shocks that are common to the entire panel—each time series has its own loading associated with the shocks. Second, the idiosyncratic components, which are specific to a particular variable and orthogonal with the past, present, and future values of the common shocks. The common component of PDs is best viewed as the result of the underlying unobserved systemic risk process, and it is thus expected that it will be relatively persistent. The idiosyncratic component instead reflects local aspects of credit risk that while far from negligible, especially in the short term, are transient.

Assume a vector of n series expressed as x_tⁱ =

α

ⁱ(L)u_t +v_tⁱ _where xt ⁼

(

xt¹,xt²,...,xtⁿ

)

^ʹ′ is a n-dimensional vector of stochastic stationary process with zero mean and variance

1;ut ⁼

(

ut¹,ut²,...,ut^q

)

^ʹ′ is a q-dimensional vector of mutually orthogonal common shocks with zero mean and unit variance, and with q<n ^; vt ⁼

(

vt^1,vt^2,,...,vt^n,

)

^{ʹ′ is a n-}

dimensional vector of idiosyncratic shocks; and

α

ⁱ(L)'^{is a (n}×^q)matrix of rational functions with the lag operator L. The model allows for correlation between vⁱ_t variables, but the variances of v_tⁱ bounded as i→∞ . When n is large, the idiosyncratic components, which are poorly correlated, will vanish, and only the common components will be left, and thus they will be identified (see Forni and others, 2000, for a technical proof).

The GDFM model is estimated using the one-sided estimator proposed by Forni et al (2005). The procedure comprises two steps: first, estimating the spectral density matrix of the vector stochastic process xⁱ_tand, second, using the calculated q largest (real) eigenvalues—and their corresponding eigenvectors—of the spectral density matrix to estimate the generalized common components. In this study, the xⁱ_t (t×n) ^vector stochastic stationary process has t = 93 monthly observations and n includes 283 market indexes and macroeconomic variables for Euro area, Belgium, Canada, Denmark, France, Germany, Greece, Japan, Netherland, Italy, Spain, Sweden, Switzerland, United Kingdom, United States, and Luxembourg. Adding the macroeconomic variables to the PDs, there are 496 (354) series for Delianedis and Geske’s (Merton’s). The number of dynamic factors is q = 3 underlying PDs or DD. Accordingly, there are 496 (354) idiosyncratic shocks. In the

α

ⁱ(L)' ⁽ⁿ×^q)matrix of rational functions with the lag operator L, the number of lags is 2, and total the number of static factors is 9.⁷

Since the common factors are derived on the standardized first difference of PDs or DD, the accumulated common component is constructed from the initial PDs or DD, and the standard deviation (STD) and mean (M) of the first difference of PDs or DD. For example, in the case of PDs, PDs_t^{AccumulatedCC} =M +

α

_i(L)u_tSTD+PDs_t^Accumulate₋₁ ^{dCC , and}

1

0 PDs

PDs^{AccumulatedCC} = . Therefore the accumulated common component shows the hypothetical evolving path of credit risk if purely driven by the common factors. The accumulated idiosyncratic component is simply the residual risk between PDs or DD and its accumulated common component. The correlation between the accumulated

7 See Hallin and Liska (2007) for the log criterion to determine the number of dynamic factors, and Alessi, Barigozzi and Capasso (2009), who modify Bai and Ng (2002) criterion for determining the number of static factors in a more robust manner.

common component and the accumulated idiosyncratic component can be statistically significant even the idiosyncratic component is orthogonal with the common factors.

3. A Dynamic Forecasting Framework

Forni et al (2005) provide a good framework for multi-step-ahead predictions of the common component. Nevertheless, the idiosyncratic (credit risk) component also plays an important role for financial instability, which cannot be neglected (see Schwaab et al, 2010). The idiosyncratic component is in general autocorrelated and therefore can be predicted. Forni et al (2003) construct a linear forecasting model with the contemporaneous common component and the lagged idiosyncratic component.

However, their forecasting method is not easily applied to a large number of underlying assets simultaneously, and also does not generate the distribution of these forecasts.

The input to the GDFM is a vector of stochastic covariance-stationary processes with zero means and finite second-order moments. However, currently there is no structural credit risk model directly combined with the GDFM. Therefore, the standardized first difference of PDs or DD (difference stationary processes) can be regarded as exogenous inputs to the GDFM. The common and idiosyncratic components are assumed to be asymptotically stationary and orthogonal to each other, whereas the idiosyncratic component can be mildly cross-correlated. Similar to the algorithms for combining GDFM and GARCH in Alessi, Barigozzi and Capasso (2007a&b), this study introduces a novel approach to combine the GDFM with a dynamic t-copula. The AR (or zero mean)-GARCH model can be applied to both the common components and the idiosyncratic components for all variables. Afterwards, a dynamic t-copula is used to glue together the standardized residuals or innovations from those marginal components.

Formally, the dynamic forecasting model becomes:

), , );

( ),..., (

), ( ( ) ,..., , (

) 1 , 0 (

~

) (

2 1 3 2

1 2 1

1 1 2

1 2

1 1

1 1

2 2 0

2 1

1 1 1

1 _

1

1 1 _

1 _

1

_ 1 _

1 1

t t n t t

t T n t t t t

t t

t t

IC t IC t p

i IC

i t F

IC t

CC t CC t F GDF t F CC t

F IC t F CC t F t

v R F

F F

C F

iid X X

X X

X X

X

+ +

+ +

+ + +

+

+ +

=

− + +

+ + +

+

+ +

+

= + +

=

+

=

+

= +

=

∑

ε ε

ε ε

ε ε ε

βσ ε

σ α α σ

ε σ

ε σ

where the forecast X_t^F₊₁of the marginal credit risk is the sum of its forecasted common component X_t^CC₊₁^_Fand idiosyncratic componentX_t^IC₊₁^_F; X_t^CC =

α

_i(L)u_t is the common

component, and X_t^IC =v_tⁱ is the idiosyncratic component from the GDFM. Both common and idiosyncratic components are simply assumed to follow a GARCH (1,1) process.

The mean of X_t^CC₊₁^_F is the prediction of the common component X_t^GDF₊₁ ^_Fby the GDFM as in Forni et al (2005), whereas the mean of X_t^IC₊₁^_F is an autoregressive process of order p, AR (p). The multivariate distributionF(

ε

¹_t+1,

ε

_t²₊₁,...,

ε

_t²₊ⁿ₁)for i=1,2,...,2n, which includes standardized residuals from both the common and the idiosyncratic components and has a time-varying t-copula form.

The copula is a fundamental tool for modeling multivariate distributions. It provides a robust method of consistent estimation for dependence, and is much flexible. Correlation, which usually refers to Pearson’s linear correlation, depends on both the marginal distributions and the copula, is not a robust measure given that a single observation can have an arbitrarily high influence on it. Instead, using the conditional dynamic copula, it is relatively easy to construct and simulate from multivariate distributions built on marginal distributions and dependence structure. Drawing on Jin and Nadal De Simone (2011a), a PD index of banking sector overall credit risk is constructed aggregating the individual banks’ PD estimates weighted by, say, their respective implied asset values.⁸ The following sections explain in detail the modelling of marginal dynamics, dynamic t- copulas, and forward simulation procedures.

3.1. Modelling Marginal Dynamics

This study does not specify marginal distributions, but adopts a semi-parametric form for the marginal distributions. Misspecification of marginal distributions can lead to dangerous biases in dependence measure estimation. This is why the semi-parametric approach is quickly becoming the standard in joint multivariate modelling. Time series data, like the common and idiosyncratic components of financial processes, usually reveal time-varying variance and heavy-tailedness. While keeping the multi-step-ahead prediction of the common components from Forni et al, 2005, a GARCH (1,1) process is fitted to the common components and an AR(p) - GARCH (1,1) process is fitted to the idiosyncratic components. The proposed marginal dynamics are formally defined as:

8 Weights other than asset values are used and discussed below.

), 1 , 0 (

~

)

( _{1 1} ^{2 2}₁

0 2

1

iid X X

X

t

t t

t t

IC t IC t p

i IC

i t IC

t

CC t CC t CC t

ε

βσ ε

σ α α σ

ε σ ε σ

−

−

−

=

−

+ +

=

+

=

=

∑

where X_t^CC is the common component, andX_t^IC is the idiosyncratic component from Forni et al (2005). The model is estimated directly by Quasi-Maximum Likelihood. The best AR (p) - GARCH (1,1) can be selected by an automatic model selection criteria, such as the Akaike Information Criterion Corrected Version (AICC). Since in the database, book-value data are actually quarterly, an AR (3) process is used to track dynamic changes, which is especially important for macroprudential policy.

Given the standardized i.i.d. residuals

ε

_tfrom the estimation of the marginal dynamics, the empirical cumulative distribution function (cdf) of these standardized residuals is estimated with a Gaussian kernel. This smoothes the cdf estimates, eliminating the rugged shape of the sample cdf. Howevr, although non-parametric kernel cdf estimates are well-suited for the interior of the distribution where most of the data are found, they tend to perform poorly when applied to the upper and lower tails. Therefore, to improve the efficiency of the tails of the distribution’s estimates, the upper and lower, e.g. 10%

thresholds of the residuals, are reserved for each tail. Then, the amount by which those extreme residuals in each tail fall beyond the associated threshold is fitted to a parametric Generalized Pareto distribution (GP) by maximum likelihood. Since in our study there are only 93 monthly observations, 20% thresholds are used to ensure that there are sufficient data points at the tails to conform well to a GP. Extreme Value Theory (EVT) in general, and in particular the GP distribution, provide an asymptotic theory of tail behavior. Under the assumption of a strict white noise process, i.e. an independent, identically distributed process, the theory shifts the focus from modelling the whole distribution to modelling tail behaviour, and hence, even asymmetry may be examined directly by estimating the left and right tails separately. In addition, EVT has the advantage of requiring just a few degrees of freedom. This approach is often referred to as the distribution of exceedances or peaks-over-threshold method (see, for instance, McNeil (1999), McNeil and Frey (2000) or Nystrom and Skoglund (2002a&b)).

3.2. The Dynamic Conditional t-Copula

As stated above, copula theory provides an easy way to deal with (otherwise) complex multivariate modeling. The main advantage of the copula approach is its flexibility. It

allows the definition of the joint distribution through the marginal distributions and the dependence between the variables. In addition, copulas are often relatively parsimoniously parameterized, which facilitates calibration. Recently, copula theory has been extended to the conditional case, allowing the use of copulas to model dynamic structures, such as in Dias and Embrechts (2004), Patton (2004, 2006a&b), and Jondeau and Rockinger (2003, 2006). The conditional copula can be a very powerful tool for active risk management as shown by Fantazzini (2009), and Jin and Lehnert (2011).

The t-copula is a good candidate for the high-dimensional problem dealt with in this paper allowing for non-zero dependence in the extreme tails. The copula of the multivariate standardized t distribution is the t-copula, and the conditional dynamic t- copula is defined as follows⁹:

)), ( ),..., ( ), ( ( )

,

; ,..., ,

( _{1 2 n t t R,v v}¹ _{1 v}¹ _{2 v}¹ _n

t t

t t

t t t t

T v R

C

η η η

= ⁻

η

⁻

η

⁻

η

where

η

_n =F_n(

ε

_n) for i=1,2,...,n, and

ε

_t ~iid(0,1),are the innovations from the marginal dynamics introduced in the previous section. R_t is the rank correlation matrix, andv_tis the degrees of freedom. _v¹( _n)

t⁻t

η

denotes the inverse of the t cumulative distribution function. R_tand v_tcan be assumed to be constant, or a dynamic process through time.

Engle (2002) proposes a class of models - the Dynamic Conditional Correlation (DCC) class of models - that preserves the ease of estimation of Bollerslev (1990)'s constant correlation model while allowing correlation to change over time. These kinds of dynamic processes can also be extended into t-copulas. The simplest rank correlation dynamics considered empirically is the symmetric scalar model where the entire rank correlation matrix is driven by two parameters:

, )

( )

1

( − − ^_+ ^*₋₁ ^*₋₁^' + ₋₁

= _{dcc t t dcc t}

t Q Q

Q

α β α ε ε β

where

α

_dcc ≥0,

β

_dcc ≥0,

α

_dcc +

β

_dcc ≤1^,

ε

_t^* =t_v⁻_t¹(

η

_n =F_n(

ε

_n))^, Q_t =|q_ij,t| is the auxiliary matrix driving the rank correlation dynamics, the nuisance parameters Q^_ =E[

ε

_t^*

ε

_t^*'| ^with

9 See Patton (2006b) for the definition of a general conditional copula.

sample analog [ ^{* *'}|

1 _ 1

t t T t

E T

Q

∑ ε ε

=

= − , so that R_t is a matrix of rank correlations q_ij,t^with

ones on the diagonal, .

, ,

, ,

t jj t ii

t ij t

ij q q

= q

ρ

Given that the correlation between the Gaussian rank correlation

ρ

_GR =Corr(Φ⁻¹(u)Φ⁻¹(v) ^{and a t-copula rank} correlation

) ( ) ( (t ¹ u t ¹ v Corr _{v v}

TR − −

ρ

= is almost equal to one, R_t can be well approximated by the

Gaussian

Rt from the dynamic Gaussian Copula. For convenience, this study adopts a two- step algorithm for estimation which means that R_t is estimated from the dynamic Gaussian copula first, and then, withR_t fixed, the degrees of freedom are recovered from the t-copula.

The dynamic multivariate Gaussian copula is defined similarly to the t-copula as follows:

)), ( ),..., ( ), ( ( )

; ,..., ,

( _{1 2 n t R}Gaussian ¹ ₁ ¹ ₂ ¹ _n

R t

C

η η η

=Φ Φ⁻

η

Φ⁻

η

Φ⁻

η

where

η

_n =F_n(

ε

_n) for i=1,2,...,n, and

ε

_t ~iid(0,1)are again the innovations from the marginal dynamics introduced in the previous section. R_t^Gaussian is the Gaussian rank correlation matrix. The rank correlation dynamics is similarly driven by the two parameters listed above for the t-copula. However,

ε

_t^* =Φ⁻¹(

η

_n =F_n(

ε

_n)).

While the quasi-likelihood function for the dynamic Gaussian copula could be computed, convergence is not guaranteed in high dimensions, and sometimes it fails or is sensitive to the starting values. This incidental parameter problem causes likelihood-based inference to have economically important biases in the estimated dynamic parameters, with specially α displaying a significant downward bias. As a result, Engle, Shephard and Sheppard (2008) suggest an approach to construct a type of composite likelihood, which is then maximized to deliver the preferred estimator:

, )

; ( 1 log

) (

1

,

1

∑

∑

= =

=

N i

t j T

t

Y N f

CL

ψ ψ

where Y_j,t is composed of all unique pairs of data, ψ is a set of parameters, N is the number of all pairs, and t=1,2,...,T. The composite likelihood is based on summing up the quasi-likelihood of all subsets. Each subset yields a valid quasi-likelihood, but this quasi-likelihood is only mildly informative about the parameters. By summing up many subsets, it is possible to construct an estimator which has the advantage of not making necessary the inversion of large dimensional covariance matrices. Further, and vitally, the estimator is not affected by the incidental parameter problem discussed above. It can also be very fast, and does not have the biases intrinsic in the usual likelihood estimator when the cross-section is large. This dynamic Gaussian copula can also be estimated by maximizing m-profile subset composite likelihood (MSCL)¹⁰ using contiguous pairs, which is attractive from statistical and computational viewpoints for large dimensional problems, at least compared with the m-profile composite likelihood (MCLE) using all the pairs. Therefore, to avoid the known estimation difficulties of high-dimensional t-copula, m-profile subset composite likelihood (MSCL) are maximized using contiguous pairs, where the degrees of freedom for the t-copula is simply the 50th quantile of all degrees of freedom derived from pairwise t-copulas.

3.3. Forward Simulation

Using conditional dynamic copulas, it is relatively easy to construct and simulate from multivariate distributions built on marginal distributions and dependence structure. The GARCH-like dynamics in both variance and rank correlation offers multi-step-ahead predictions of the common and the idiosyncratic components simultaneously.

The following steps illustrate the one-step-ahead simulation:

1. Draw independently

ε

_t^*₊ⁱ¹₁,...,

ε

_t^*₊^im₁ for each component from the n-dimensional t distribution with zero mean, forecast correlation matrix R_t+1, and degrees of freedom v_t+1 to obtain

µ

_tⁱ¹₊₁,...,

µ

_t^im₊₁ by setting ₁ ( ^*₁)

1

ik t v ik

t₊ t _{t +}

= +

ε

µ

, where k=1,...,m, the total paths of simulation, i=1,...,n, the number of components;

2. Obtain

ε

_tⁱ¹₊₁,...,

ε

_t^im₊₁by setting

ε

_t^ik₊₁=F_i⁻¹(

µ

_t^ik₊₁)^{, where}Fi is the empirical marginal dynamics distribution for component i;

3. Obtainz_tⁱ¹₊₁,...,z_t^im₊₁by setting z_t^ik₊₁=

ε

_t^ik₊₁

σ

_tⁱ₊₁^{, where}

σ

_tⁱ₊₁is the forecast standard deviation using a GARCH (1,1) model for component i;

10 A moment-based profile likelihood, or m-profile likelihood for short, in which the nuisance parameters are not maximum quasi-likelihood estimators but attractive moment estimators.