Journal of Network Theory in Finance

Risk.net

Relation between regional uncertainty spillovers in the global banking system

Sachapon Tungsong, Fabio Caccioli and Tomaso Aste

  • We add exponential weighting to the method of Diebold and Yilmaz.
  • We construct time series of uncertainty for three regional banking systems.
  • We find significant causal relations between uncertainty in the three regions.
     

We report on time-varying network connectedness within three banking systems: North America (NA), the European Union (EU) and Southeast Asia (ASEAN). Diebold and Yilmaz’s original method is improved by using exponentially weighted daily returns as well as ridge regularization on vector autoregression (VAR) and fore- cast error variance decomposition (FEVD). We compute the total network connectedness for each of the three banking systems, which quantifies regional uncertainty. Results over rolling windows of 300 days during the period from January 2005 to October 2015 reveal changing uncertainty patterns that are similar across regions, demonstrating common peaks associated with identifiable exogenous events. Lead– lag relationships among changes of total network connectedness of the three systems, quantified by transfer entropy, reveal that uncertainties in the three regional systems are significantly causally related, with the NA system having the largest influence on the EU and ASEAN.

1 Introduction

Financial markets are increasingly becoming more interconnected (Moghadam and Vinals 2010), and shocks initially affecting one part of the system can quickly propagate to the rest of it. Therefore, understanding the patterns of distress propagation within financial markets is important to characterize systemic risk. After the global financial crisis of 2007–9, significant effort has been devoted to understanding the mechanics of distress propagation within banking systems. One strand of the literature has focused on modeling the processes through which contagion may occur in interbank networks (see, for example, Glasserman and Young (2016) and Caccioli et al (2018) for recent reviews). Another strand of the literature has focused on the quantification of systemic risk from market data (see Adrian and Brunnermeier 2016; Brownlees and Engle 2016). In particular, Diebold and Yilmaz (2009) proposed a method based on forecast error variance decomposition (FEVD) of estimating from market data networks of interdependencies between firms, and they used the connectedness of the estimated networks to quantify spillovers of uncertainty between variables.

In this paper, we use the methodology of the aforementioned work by Diebold and Yilmaz (2009) to estimate the time evolution of connectedness in three regional banking systems: North America (NA), the European Union (EU) and Southeast Asia (ASEAN). Through VAR and FEVD, we compute the pairwise connectedness between pairs of banks in each region, and we aggregate such pairwise connectedness to compute a measure of total connectedness for the region.

The time-varying total connectedness computed for each banking system, from a 300-day rolling window during the period from January 2005 to October 2015, indicates temporal changes of systemic risk, with peaks during major crisis events and troughs during normal periods. Analogous results have been observed in other financial systems and different regions (Diebold and Yilmaz 2009, 2012, 2014; Chau and Deesomsak 2014; Alter and Beyer 2014; Fengler and Gisler 2015; Demirer et al 2018). It has to be stressed that, unlike Diebold and Yilmaz (2009), who view all financial institutions as belonging to one global system, here we group banks into three regional banking systems. In this way, we can perform a comparative analysis between the different regions, which allows us to highlight similarities and differences between them. Further, this allow us to quantify the existence of causal relations between different regions. Note that combining all the banks together could be somehow misleading, because the banks’ equities in the three banking systems trade in different stock markets that have significantly different trading hours.

The main results of our analysis are as follows. First, we note that the structure of the peaks in the three regional banking systems is very similar, with large peaks associated with significant, identifiable major events. Although the overall patterns are similar, we observe two important differences between the systems. The first is the fact that the overall scale of connectedness is different, with the NA banking system being more interconnected than that in the EU, which, in turn, is more interconnected than the ASEAN system. Second, we uncover the existence of lead–lagged relations between the different time series. To quantify this effect, we compute the transfer entropy between the time series associated with changes of connectedness in the different regions, and we uncover the existence of significant net information flows from NA to the EU, from NA to ASEAN, and from the EU to ASEAN. The robustness of our finding is tested using different measures for transfer entropy. In particular, we find consistent results for the net information flow both with a linear measure of transfer entropy (which corresponds to a Granger causality analysis) and with nonlinear measures of different parameters. We also retrieve similar causal relations for both one-day and five-day returns. To the best of our knowledge, this causality study between regional uncertainties is the first of its kind.

The rest of this paper is organized as follows. In Section 2, we present a literature review and place our paper in the context of previous works. In Section 3, we describe the used data, while Section 4 provides a brief description of our methodology. Section 5 illustrates and discusses the main results of the paper. We present our conclusions in Section 6.

2 Literature review

The literature on systemic risk and contagion in the banking network can be broadly classified into two categories. The first category comprises network models that aim to describe various causal mechanics of financial contagion, which can be calibrated with balance-sheet data (Furfine 2003; Degryse and Nguyen 2007; Upper and Worms 2004; Müller 2006; Cont et al 2010; Upper 2011; Birch and Aste 2014). The second category comprises econometric models that aim to identify spillover effects exclusively from market data, without making assumptions about the dynamics of distress propagation between banks (Adrian and Brunnermeier 2016; Brownlees and Engle 2016). Our paper is close to this second strand of the literature, as we try to understand whether market data carries information about the level of interconnectedness between banks and how exogenous shocks can be amplified by the endogenous dynamics of financial markets.

Network models of contagion go back to the seminal work of Allen and Gale (2000), who showed how the stability of the banking system is affected at equilibrium by the pattern of interconnections between banks, and to the work of Eisenberg and Noe (2001), who demonstrated how to consistently compute a clearing vector of payments in a network of interbank claims. The relation between the structure of an interbank network and its stability has also been extensively explored within the context of nonequilibrium network models (see, for example, Furfine 2003; Iori et al 2006; Nier et al 2007; Gai and Kapadia 2010; Cont et al 2010; Upper 2011; Battiston et al 2012; Fricke and Lux 2015; Bardoscia et al 2015, 2017; Kobayashi and Hasui 2014; Lenzu and Tedeschi 2012; Tedeschi et al 2001), showing in particular the existence of a tension between individual risk and systemic risk: what makes a bank individually less risky might in fact increase the risk of a systemic failure (Beale et al 2011). More recently, these analyses have been extended beyond interbank lending networks to the study of networks of overlapping portfolios (Huang et al 2013; Caccioli et al 2014; Corsi et al 2016). Although these models have been insightful with regard to understanding the dynamics of financial contagion, and in some cases they have been applied to real data (see Upper (2011) for a review of the existing literature), there are clear challenges to their applicability. First, there is a lack of reliable data on banks’ balance sheets, which makes it hard to calibrate models.11Admati et al (2013) report that banks tend to find ways to get around regulations in order to invest in mortgage-backed securities and derivatives via structured-investment vehicles, which are off-balance-sheet items. Such leeway being allowed by regulations creates regulatory boundaries, making it difficult for outsiders to know what banks actually report. Second, to obtain a reliable assessment of systemic risk one has to capture all relevant types of interconnections between banks, as the interaction between different contagion channels can significantly change the stability of the system (Caccioli et al 2015).

Here, we take the complementary approach of inferring interdependencies between banks from market data, which belongs to the second strand of literature mentioned above. The advantage of this approach with respect to network modeling is that market data is readily available, and different types of interconnections between banks have already been aggregated by the market. The drawback is that this approach does not provide an explanation of how stress propagates between banks, and it relies on the underlying assumption of market efficiency, which is not realistic (Shiller 2003). Nevertheless, one can assume that, although markets are not efficient, prices do reflect to some extent the aggregate information (or expectations) about the underlying assets. There have been several contributions to this strand of the literature. In particular, Dungey et al (2005) provide a summary of empirical models of contagion up to 2005. More recent empirical work includes Diebold and Yilmaz (2009, 2012, 2014), Caceres et al (2010), Billio et al (2012), Claeys and Vasicek (2014), Lucas et al (2014), Musmeci et al (2015) and Brownlees and Engle (2016). Of particular relevance for our paper is the work of Diebold and Yilmaz (2009, 2012, 2014), which influenced subsequent studies such as McMillan and Speight (2010), Bubák et al (2011), Fujiwara and Takahashi (2012), Klößner and Wagner (2014), Alter and Beyer (2014), Chau and Deesomsak (2014), Demirer et al (2018) and Fengler and Gisler (2015). This strand of contributions uses vector autoregression (VAR) and forecast error variance decomposition (FEVD) to quantify the unpredictability of each variable in the network. By using the VAR and FEVD methods, it is possible to disentangle the contribution to unpredictability due to endogenous interdependencies from that due to exogenous shocks. Following Diebold and Yilmaz, we will refer to this endogenous component in our paper as total network connectedness, which therefore quantifies the transmission of shocks from banks within the system.

3 Data

We collect daily stock prices from January 2005 to October 2015 of banks headquartered in NA (including the United States and Canada), the EU and ASEAN from the Compustat database. We select only financial institutions in the subindustry “Banks” (ie, large banks operating at the national level and having a GICS code of 40101010) and compute log returns from the daily closing prices for each bank. With the aforementioned criteria, our sample includes ten publicly listed banks in NA, sixty-six banks in the EU and thirty-nine banks in ASEAN that survived through the period January 2005–October 2015.

While we could analyze rolling windows in which the number of banks in operation varies from one window to the next, we find that being able to see the evolution of the systems’ total connectedness given a constant number of banks provides some baseline insight into how the same set of banks reacted to different economic and financial episodes over time. That being said, research that saw all surviving banks being accounted for in respective rolling windows would be an interesting avenue to explore. In such a case, the dimensions of the rolling windows would likely be much larger and estimation techniques such as sparsity modeling would be needed.

All banks in the NA banking system have their stocks traded in the New York Stock Exchange (NYSE), while the EU and ASEAN bank stocks mostly trade in their own national stock markets. Lists of banks in all three regions as well as their summary statistics are given in Tables 13.

Table 1: List of banks headquartered in NA (Canada and the United States) that have actively traded between 2005 and 2015.
    Daily Daily
    mean volatility
Bank name Country return (%) (%)
Canadian Imperial Bank (CIBC) CAN -0.01 1.82
Bank of Montreal (BMO) CAN -0.01 1.69
Royal Bank of Canada (RBC) CAN -0.03 1.73
Toronto Dominion Bank (TD) CAN -0.03 1.65
Bank of Nova Scotia (BNS) CAN -0.01 1.72
Citigroup (CITI) USA -0.08 3.70
Bank of America Corp (BAC) USA -0.04 3.51
Wells Fargo & Co (WFC) USA -0.02 2.86
JP Morgan Chase & Co (JPM) USA -0.02 2.64
US Bancorp (USB) USA -0.01 2.32
Table 2: List of banks headquartered in ASEAN that have actively traded between 2005 and 2015.
    Market cap Average Volatility
Bank Country (USD billion) return (%) (%)
Bank Rakyat Indonesia IDN 20.43 0.07 2.56
Bank Permata IDN 00.54 0.02 1.93
Bank Danamon IDN 02.23 0.00 2.73
Bank Maybank Indonesia IDN 00.79 0.00 2.67
Bank Cimb Niaga IDN 01.07 0.02 2.51
Panin Bank IDN 00.17 0.03 2.68
Bank Negara Indonesia IDN 06.66 0.04 2.50
Bank Central Asia IDN 23.21 0.08 2.06
Bank Mandiri IDN 15.75 0.05 2.54
Public Bank MYS 16.15 0.04 0.90
Malayan Banking MYS 18.70 0.00 1.23
RHB Capital MYS 03.73 0.03 1.58
AMMB Holdings MYS 03.04 0.01 1.51
AFFIN Holdings MYS 00.97 0.01 1.65
Alliance Financial Group MYS 01.15 0.01 1.52
BIMB Holdings MYS 01.35 0.03 2.13
CIMB Group Holdings MYS 07.92 0.02 1.54
Hong Leong Bank MYS 06.17 0.03 1.14
Philippine National Bank PHL 01.20 0.03 2.39
Bank of Philippine Islands PHL 06.97 0.03 1.79
China Banking Corp PHL 01.36 0.04 1.39
Metropolitan Bank and Trust PHL 04.67 0.05 2.12
Security Bank Corp PHL 01.86 0.07 1.87
Rizal Commercial Bank Corp PHL 00.94 0.03 2.19
Union Bank PHL 01.22 0.05 1.77
BDO Unibank PHL 07.33 0.05 2.04
United Overseas Bank SGP 19.62 0.01 1.49
DBS Group Holdings SGP 25.23 0.01 1.49
Oversea-Chinese Banking SGP 22.71 0.02 1.33
Krung Thai Bank THA 06.79 0.02 2.11
Siam Commercial Bank THA 11.44 0.03 2.02
Bangkok Bank THA 08.04 0.02 1.81
Bank of Ayudhya THA 06.15 0.03 2.41
Kasikornbank THA 10.94 0.04 1.97
TMB Bank THA 03.12 -0.01 2.40
Kiatnakin Bank THA 00.91 0.00 1.94
Tisco Financial Group THA 00.96 0.02 2.11
Thanachart Capital THA 14.3 0.03 2.13
CIMB Thai Bank THA 00.76 -0.01 2.75
Table 3: List of banks headquartered in the EU that have actively traded between 2005 and 2015.
    Daily Volatility
Bank Country return (%) (%)
Oberbank Ag AUT -0.02 0.38
Erste Group Bk Ag AUT -0.01 2.95
KBC Group Nv BEL -0.00 3.50
Dexia Sa BEL -0.21 7.76
Hellenic Bank CYP -0.08 3.08
Komercni Banka As CZE -0.01 2.10
IKB Deutsche Industriebank DEU -0.13 3.90
Commerzbank DEU -0.08 3.09
DVB Bank Ag DEU -0.03 1.38
HSBC Trinkaus & Burkhardt DEU -0.00 1.73
Comdirect Bank Ag DEU -0.02 1.83
Deutsche Postbank Ag DEU -0.00 2.15
Danske Bank As DNK -0.01 2.11
Jyske Bank DNK -0.02 1.94
Nordea Invest Fjernosten DNK -0.01 1.43
Sydbank As DNK -0.03 1.93
Banco Santander Sa ESP -0.00 2.16
BBVA ESP -0.01 2.12
Banco Popular Espanol ESP -0.07 2.30
Bankinter ESP -0.01 2.28
Banco De Sabadell Sa ESP -0.02 1.89
BNP Paribas FRA -0.00 2.56
Natixis FRA -0.01 3.12
Societe Generale Group FRA -0.02 2.86
Credit Agricole Sa FRA -0.02 2.78
CIC (Credit Industriel Comm) FRA -0.00 1.41
Barclays Plc GBR -0.03 3.23
HSBC Hldgs Plc GBR -0.02 1.72
Royal Bank of Scotland Group GBR -0.10 3.91
Standard Chartered Plc GBR -0.00 2.44
Lloyds Banking Group Plc GBR -0.05 3.37
Piraeus Bank Sa GRC -0.22 5.04
Attica Bank Sa GRC -0.23 5.88
Eurobank Ergasias Sa GRC -0.31 5.52
National Bank of Greece GRC -0.20 4.81
Alpha Bank Sa GRC -0.15 4.69
Zagrebacka Banka HRV -0.00 2.58
Privredna Banka Zagreb Dd HRV -0.01 2.37
OTP Bank Plc HUN -0.00 2.63
Unicredit Spa ITA -0.05 2.90
Credito Emiliano Spa ITA -0.00 2.26
Intesa Sanpaolo Spa ITA -0.00 2.61
Banca Popolare Di Sondrio ITA -0.01 1.83
Banca Carige Spa Gen & Imper ITA -0.10 2.39
Banco Desio Della Brianza ITA -0.02 1.76
Banco Popolare ITA -0.06 2.86
Banca Popolare Di Milano ITA -0.03 2.78
Banca Monte Dei Paschi Siena ITA -0.12 2.96
Bank of Siauliai Ab LTU -0.06 2.97
ING Groep Nv NLD -0.01 3.14
Van Lanschot Nv NLD -0.03 1.62
Mbank Sa POL -0.05 2.34
Bank Handlowy W Warzawie Sa POL -0.01 2.05
ING Bank Slaski Sa POL -0.04 1.90
Bank BPH Sa POL -0.09 4.48
Bank Millennium Sa POL -0.03 2.62
Bank Plsk Kasa Opk Grp Pekao POL -0.00 2.26
Bank Zachodni Wbk Sa POL -0.04 2.15
Getin Holding Sa POL -0.02 3.16
Powszechna Kasa Oszczednosci POL -0.00 2.02
Banco BPI Sa PRT -0.03 2.46
Banco Comercial Portugues Sa PRT -0.09 2.76
Svenska Handelsbanken SWE -0.02 1.86
Skandinaviska Enskilda Bank SWE -0.01 2.55
Nordea Bank Ab SWE -0.02 2.05
Swedbank Ab SWE -0.01 2.53

The data was analyzed over rolling windows of 300 days and over the full period. Harris (1985) recommends using a sample size such that n50+k, where k is the number of predictors. For our study, the minimum number of observations for each rolling window is thus 50+63=113. We experimented with window sizes of 250, 500 and 750 days and obtained similar results in terms of the overall shape, including peaks and troughs, of total connectedness. We chose the window size of 300 days because it is a good compromise between obtaining results with a reasonable margin of error and making sure we cover the period of interest (March 2006 to November 2015). A window size of 500 days would provide results with a lower margin of error but cover the period from January 2007 onward, while a window size of 250 days would provide results with a higher margin of error but cover the period from January 2006 onward.

4 Methodology

4.1 Total connectedness

Following the approach introduced by Diebold and Yilmaz (2009, 2012, 2014), we use a variance decomposition whereby the forecast error variance of a variable is decomposed into contributions attributed to each variable in the system. The approach is based on the VAR model, introduced by Sims (1980) (see Stock and Watson (2001), Cochrane (2005), Lutkepohl (2006) and Tsay (2010) for discussions, reviews and applications).

VAR estimates the value of a set of N variables yt,1,,yt,N at time t from a linear combination of their values in the past by performing a multidimensional regression. By using the vectorial representation ?t=(yt,1,,yt,N)T and considering the t-1 lag only, the regression can be written as ?t=??t-1+?t, with ? an N×N matrix of coefficients. By iterating this formula and expressing it in terms of an orthonormal basis of residuals wi,t (with var(wi,twj,t)=δi,j) (Cochrane 2005), one can write

  yi,t=s=0j=1Nθij,swj,t-s.   (4.1)

The one-step-ahead forecast is ?^t+1=??t. The forecast error is the difference yi,t+1-y^i,t+1=θij,0wj,t+1, and its variance is therefore

  var(yi,t+1-y^i,t+1)=j,k=1Nθij,0θik,0var(wj,t+1,wk,t+1)=j=1Nθij,02.   (4.2)

Each term θij,02 in the sum is interpreted as the contribution to the one-step forecast error variance of variable i due to shocks in variable j. Its normalized value, cij=θij,02/k=1Nθik,02, is called connectedness by Diebold and Yilmaz (2009, 2012, 2014) and it is associated with the relative uncertainty spillover from variable j to variable i. In this paper, we will report on the “total connectedness”, which is

  total connectedness=1Ni,j=1ijNcij   (4.3)

and measures the average effect that the variables have on the one-step forecast error variance. It is a measure of spillover uncertainty across the entire system. Larger values of total connectedness correspond to unstable periods in which variables’ uncertainties strongly influence one other.

We refine the original Diebold and Yilmaz (2009, 2012, 2014) methodology by introducing two technical improvements. The first improvement consists in employing ridge regularized VAR (Tikhonov 1963; Hoerl and Kennard 1970), which is used to make estimations less sensitive to the noise and uncertainty associated with having a time series of finite length. Ridge regression introduces a penalty on the square sum of regression coefficients, thus favoring models with smaller coefficients. This improves regression performances, especially for systems with a large number of variables, where the covariance matrix is nearly singular (see Gruber 1998). In practice, ridge regression consists in adding a diagonal term in the expression for the regression coefficients: B=(XX+λI)-1XY, with I the identity matrix and λ a coefficient that makes the inversion less sensitive to uncertainty over small eigenvalues (Tikhonov 1963). The parameter λ must be chosen with respect to regression performances; it depends on the length of the time series and on their statistical properties. In our case, we used λ=100, which we determined was a good compromise value for this data set, and a window length of 300 points.22We multiplied returns by a factor 100 in our analysis. Therefore, the value λ=100 is reasonable compared with the norm of the matrix XX, which is of order 104. We verified that the results are slightly sensitive to variations of λ in a wide range [1001000]. The second technical improvement consists in using exponential smoothing to mitigate the effects associated with sensitivity to large variations in remote observations (Pozzi et al 2012). Exponential smoothing computes weighted averages over the observation window, with exponentially decreasing weights, exp(-s/θ), assigned to more remote observations (here, s counts the number of points from the present). In this paper, we use rolling windows of size 300 days with exponential weights of characteristic length θ=100. Choosing a characteristic length approximately one-third of the window’s length was suggested as optimal by Pozzi et al (2012).

4.2 Transfer entropy and Granger causality

We investigate how uncertainty in one region affects uncertainty in another region by quantifying lead–lag relationships among uncertainty spillovers. For this purpose, we compute the transfer entropy associated with the daily and weekly changes in the total connectedness of the three systems.

In this paper, we estimate the transfer entropy by using both linear and nonlinear approaches. The transfer entropy TYX quantifies the reduction of uncertainty on the variable X that is provided by the knowledge of the past of the variable Y taking into consideration the information from the past of X. In terms of conditional entropies, it can be written as

  TYX=H(XtXt-lag)-H(XtXt-lag,Yt-lag),   (4.4)

where Xt represents the present value of variable X and Xt-lag its lagged past. In this paper, we report the results for one-day lag. The conditional entropies are defined as H(AB)=H(A,B)-H(B), with H(A,B) the joint entropy of variables A and B and H(B) the entropy of variable B.

For what concerns the computation of these entropies, the linear approach is the standard procedure. It quantifies the additional reduction in the variance of a variable Y provided by the past of variable X, and it is directly related to Granger causality (Granger 1988; Barnett et al 2009). In this linear case, the entropy associated with a set of variables Z is proportional to the log determinant of the covariance H(Z)=12logdet(2eπΣ(Z)), where Σ(Z) is the covariance matrix of the variables in Z. The result of using (4.4) is that TYX is simply given by half the logarithm of the ratio between the regression error of variable X regressed with respect to Xt-lag and the regression error of variable X regressed with respect to both Xt-lag and Yt-lag. The nonlinear approach instead estimates entropies by first discretizing the signal into three states, associated with a central band of values within δ standard deviations from the mean and two external bands, respectively, with values smaller or larger than the central band. By calling pA0, pA- and pA+ the relative frequencies of the observations in the three bands, entropy is estimated as H(A)=-pA-logpA--pA0logpA0-pA+logpA+. The joint entropies are equivalently defined by the joint combination of values of the variables in the three bands, and the transfer entropy is retrieved by applying (4.4).

The information flow can be measured by comparing transfer entropies in both directions. If TYX>TXY, then one can say that the direction of the information is prevalently from Y to X; conversely, if TXY>TYX, then the direction of the information is prevalently from X to Y. The net information flow between X and Y can be quantified as TXY-TYX.

We validated the statistical significance of transfer entropy by comparing our results with the null hypothesis generated by computing 10 000 values of the transfer entropy, which in turn was obtained by randomizing the order of the lagged variables. This provides a nonparametric null hypothesis from which p-values can be computed. We also compared this nonparametric p-value estimate with the one from F-statistics in the linear case and found comparable results.

5 Results

5.1 Total connectedness

Using data from January 2005 to October 2015, we compute the total connectedness of the three banking systems – NA, the EU and ASEAN – over a rolling window of 300 days for the ten-year period from March 2006 to November 2015. Figures 1, 2 and 3 report the results for each of the three systems, comparing the original approach of Diebold and Yilmaz (2009) (dashed red lines) with the improved approach proposed in this paper (solid blue lines). Let us first observe that the two approaches demonstrate similar values and comparable behavior with regard to total connectedness. We can see that the use of ridge regularized VAR eliminates some of the outlying spurious peaks observed with the original method. The effect is present in all samples across the three regions and periods, but it is particularly evident in Figure 2 for the peaks after January 2011 and January 2012. When dimensionality is high, as in the case of the EU banking system, ordinary least squares estimates tend to have high variance as a result of overfitting. Ridge regression provides parameter estimates that have low variance across rolling windows, which is a manifestation of the model’s ability to better generalize across different samples. This is why we observe no sudden jumps in the total connectedness when we estimate our VAR coefficients using ridge regression. More evident is the effect of exponential smoothing, which makes peaks sharper and eliminates the plateau effect due to the persistence of the influence of a peak during the whole length of the rolling window. This is especially evident in Figure 1, where for the standard VAR method the peak in total connectedness observed just after January 2009 persists, creating a plateau that drops abruptly after 300 days in January 2010. Conversely, the exponential weighted ridge regularized method reveals a clear peak, reaching its maximum around January 2009, followed by a sharp decrease. We observe that the plateau effects in the standard VAR-equal-weights method sometimes completely hide peaks that are instead detected with the exponentially weighted ridge regularized method. This is the case for the late-2010 NA spillover peak, which is visible in Figure 3 only for the exponentially weighted ridge regularized method.

Note that in Diebold and Yilmaz (2009), where total connectedness in both equity index returns and equity index return volatilities was measured, the authors found that the return spillovers demonstrated “a gently increasing trend but no bursts, whereas volatility spillovers display[ed] no trend but clear bursts”. Our results in Figures 1, 2 and 3 indicate that applying exponential weights to the returns allows us to observe both trends and bursts in the return uncertainty spillovers.

ASEAN banking system: comparison between total connectedness computed with classical VAR approach (dashed red line) and the proposed approach (solid blue line), with ridge penalization and exponential smoothing. Computations are over a 300-day rolling window.
Figure 1: ASEAN banking system: comparison between total connectedness computed with classical VAR approach (dashed red line) and the proposed approach (solid blue line), with ridge penalization and exponential smoothing. Computations are over a 300-day rolling window.
EU banking system: comparison between total connectedness computed with classical VAR approach (dashed red line) and the proposed approach (solid blue line), with ridge penalization and exponential smoothing. Computations are over a 300-day rolling window.
Figure 2: EU banking system: comparison between total connectedness computed with classical VAR approach (dashed red line) and the proposed approach (solid blue line), with ridge penalization and exponential smoothing. Computations are over a 300-day rolling window.
NA banking system: comparison between total connectedness computed with classical VAR approach (dashed red line) and the proposed approach (solid blue line), with ridge penalization and exponential smoothing. Computations are over a 300-day rolling window.
Figure 3: NA banking system: comparison between total connectedness computed with classical VAR approach (dashed red line) and the proposed approach (solid blue line), with ridge penalization and exponential smoothing. Computations are over a 300-day rolling window.

A comparison between ASEAN, EU and NA total connectedness from the ridge regularized VAR models is presented in Figure 4, where major events are labeled on the graph when they occurred. The general shapes of the total connectedness of the three banking systems appear to be similar. Over the approximately ten-year period from March 2006 to November 2015, the values of NA’s total connectedness are generally higher than those of the EU and ASEAN banking systems, except in the following periods: 2006 to mid-2007, early 2011, early 2013 and mid-2014.

The fact that NA, EU and ASEAN banking systems have different levels of interconnectivity reflects the dissimilarities in the natures of the three banking systems. Our data set includes large banks operating at the national level (GICS code 40101010) that survived in the period from January 2005 to October 2015. Based on the GICS code and survival criteria, our NA system covers two countries (ten banks), the EU covers seventeen countries (sixty-six banks) and ASEAN covers five countries (thirty-nine banks). The two countries in the NA system (the United States and Canada) have banking regulations that are more similar than those of the seventeen countries in the EU or those of the five countries in ASEAN. In addition, the equities of the ten banks in NA trade on the same stock exchange – the NYSE – while those of the EU and ASEAN banks trade on different national stock exchanges. Finally, as banks tend to form business relationships with other banks that are in close proximity, both geographically and from a regulatory perspective, the number of interbank business activities in NA is likely to be higher than in the EU and ASEAN. These three factors contribute to stronger links and a higher possibility of spillovers among NA banks than among EU or ASEAN banks. For the above reasons, total connectedness in the NA system is generally higher than in the EU and ASEAN systems.

The number of banks in a system does not seem to be a factor that influences the level of total connectedness, as there is no relationship between the number of banks and total connectedness in a system. Note that the total connectedness metric is computed on a per-bank basis; it is the average of all pairwise connectedness in a system.

From visual inspection of Figure 4, we note that variations in total connectedness of the NA banking system seem to lead those of the EU and ASEAN systems, while the total connectedness of the EU system seems to lead that of the ASEAN system. This prompts us to perform causality tests on the total connectedness time series of the three banking systems in order to investigate how systemic uncertainty in each region influences the others as well as the lead–lag relationships among them.

Total connectedness in the three banking systems (as in Figures ...; solid lines). Major events associated with peaks are indicated by letters in the figure: A, subprime mortgage crisis; B, securitization market closedown; C, global stock market sharp fall; D, US near-record deficit USD410 billion; E, nationalization of Northern Rock; F, Fannie Mae/Freddie Mac rescue; G, Lehman Brothers filed for bankruptcy; H, rescue of RBS, Lloyds and HBOS; I, IMF approved US\$2.1 billion loan for Iceland; J, US government gave Bank of America USD20 billion aid; K, RBS reported £2.1 billion loss; L, 12.5% economic contraction in Japan; M, Greek 120 billion euro bailout; N, US credit downgrade from A+ to A; O, London Interbank Offered Rate (Libor) scandal; P, City of Detroit bankruptcy; Q, Ukranian/Syrian/Egypt unrest; R, Ebola epidemic. Computations are over a 300-day rolling window.
Figure 4: Total connectedness in the three banking systems (as in Figures 1–3; solid lines). Major events associated with peaks are indicated by letters in the figure: A, subprime mortgage crisis; B, securitization market closedown; C, global stock market sharp fall; D, US near-record deficit USD410 billion; E, nationalization of Northern Rock; F, Fannie Mae/Freddie Mac rescue; G, Lehman Brothers filed for bankruptcy; H, rescue of RBS, Lloyds and HBOS; I, IMF approved USD2.1 billion loan for Iceland; J, US government gave Bank of America USD20 billion aid; K, RBS reported £2.1 billion loss; L, 12.5% economic contraction in Japan; M, Greek €120 billion bailout; N, US credit downgrade from A+ to A; O, London Interbank Offered Rate (Libor) scandal; P, City of Detroit bankruptcy; Q, Ukranian/Syrian/Egypt unrest; R, Ebola epidemic. Computations are over a 300-day rolling window.
Table 4: Quantification of transfer entropy between regional total connectedness (March 28, 2006–November 2, 2015; full sample): from daily changes in the total connectivity using a one-day lag. (*p-value<0.05. **p-value<0.01. ***p-value<0.001.)
      Net
      information
Method ?????? ?????? flow
Linear 0.004722** 0.001354* 0.003369
Nonlinear threshold σ 0.005251*** 0.006711** -0.001460
Nonlinear threshold 2σ 0.003980*** 0.002012* 0.001968
Nonlinear threshold 3σ 0.004939*** 0.000561 0.004378
      Net
      information
Method ?????? ?????? flow
Linear 0.017336*** 0.008931*** 0.008405
Nonlinear threshold σ 0.008789*** 0.005837** 0.002953
Nonlinear threshold 2σ 0.005348*** 0.002305* 0.003042
Nonlinear threshold 3σ 0.003150** 0.002803*** 0.000348
      Net
      information
Method ?????? ?????? flow
Linear 0.005659** 0.003633** 0.002026
Nonlinear threshold σ 0.005553** 0.001262 0.004291
Nonlinear threshold 2σ 0.005960*** 0.000228 0.005732
Nonlinear threshold 3σ 0.004238*** 0.002118*** 0.002120
Table 5: Quantification of transfer entropy between regional total connectedness (March 28, 2006–November 2, 2015; full sample): from weekly changes (five days) in the total connectivity using a five-day lag. (*p-value<0.05. **p-value<0.01. ***p-value<0.001.)
      Net
      information
Method ?????? ?????? flow
Linear 0.008003* 0.001255 0.006747
Nonlinear threshold σ 0.009204 0.009474 -0.000271
Nonlinear threshold 2σ 0.017228*** 0.003196 0.014032
Nonlinear threshold 3σ 0.024087*** 0.002335* 0.021752
      Net
      information
Method ?????? ?????? flow
Linear 0.017200** 0.003703 0.013497
Nonlinear threshold σ 0.010598* 0.004354 0.006244
Nonlinear threshold 2σ 0.006509 0.006475 0.000034
Nonlinear threshold 3σ 0.002107 0.006805*** -0.004698
      Net
      information
Method ?????? ?????? flow
Linear 0.022020** 0.000619 0.021401
Nonlinear threshold σ 0.021641*** 0.002374 0.019267
Nonlinear threshold 2σ 0.022964*** 0.002900 0.020063
Nonlinear threshold 3σ 0.007488** 0.000405 0.007083

5.2 Causality tests on regional total connectedness

In order to quantify the lead–lag relationships among the NA’s, EU’s and ASEAN’s total connectedness, we compute transfer entropy and information flow between the daily changes of total connectedness in the three regions for a one-day lag. Results are reported in Table 4. Transfer entropies are estimated using both the linear and the nonlinear approaches discussed in Section 4.2. We recall that the linear measure is equivalent to Granger causality, where a significant transfer entropy corresponds to a validated Granger causality relation. The nonlinear measures are computed for fluctuation bands at δ=1,2,3 standard deviations (see Section 4.2). Observe that there is a significant information transfer between NA and EU, NA and ASEAN, and EU and ASEAN that, for the linear case, implies NA Granger causes EU, NA Granger causes ASEAN, and EU Granger causes ASEAN. We observe that the nonlinear estimation gives consistent results with the linear estimate for all values of δ, demonstrating the robustness of the result. We also observe that there are significant causal relations in the opposite direction. Given the extended time-lags between the three regions, it is fair to question whether a one-day time lag and a one-day time horizon will affect markets asymmetrically depending on their relative opening hours. We therefore test the flow of information across regions for a time horizon and lag of five days instead of one day. The results for the transfer entropies and information flow, performed for the entire period on nonoverlapping five-day returns, are reported in Table 5. We observe that the results are consistent with those for a one-day time horizon and lag reported in Table 4, the main difference being the lower statistical significance. This is expected because the time series for the five-day changes are five times shorter than those for daily changes.

6 Conclusion

We investigate regional and inter-regional uncertainty spillovers in the NA, EU and ASEAN banking systems during a period characterized by great regional and global financial stress (2005–15). Uncertainty and financial instability is quantified by means of total network connectedness, which we measure by improving on the method of Diebold and Yilmaz (2009). We demonstrate that exponential smoothing and ridge regression provide better-defined peaks in the temporal analysis and avoid the occurrence of some spurious peaks. We observe that the NA system appears to be consistently more interconnected than that of the EU, which in turn is more interconnected than the ASEAN network. Similarly to the previous analysis of Diebold and Yilmaz of other systems, our empirical analysis of the NA, EU and ASEAN banking networks shows that increased connectivity corresponds to periods of higher distress in the system. We observe that all large peaks of total network connectedness are associated with identifiable major exogenous events. Despite some of these events being related to specific regions, the effects are seen across all three banking systems, which reveal similar patterns of peaks and troughs in the variations of their total network connectedness. However, such variations are not perfectly synchronous across the regions, and causality patterns are discovered using transfer entropy. Our analysis reveals that the NA banking system is the most influential, having the largest effect on the other systems. However, feedback effects are measured with significant causal relations in the opposite direction as well. The results are demonstrated to be robust with respect to changes in the method used to compute the transfer entropy, changes in the values of parameters, and with respect to the use of daily or weekly returns in the analysis.

To summarize, the contribution of this paper is threefold. First, we improve the technical aspect of VAR estimation, allowing for better identification of events concentrated at specific times, which leads to a more accurate and insightful interpretation of the results. Second, we focus on connectedness in the banking sector, while previous studies based on the Diebold and Yilmaz (2009) methodology have analyzed networks of financial institutions. In particular, we analyze the NA, EU and ASEAN banking systems individually and show that, despite the regions’ being geographically distant, they are affected to varying degrees by major financial crisis events originating in dominant regions such as the NA and EU banking systems. Third, we perform a causality analysis on the regional connectedness time series generated using Diebold and Yilmaz’s method. Our analysis suggests that a regional disaggregated investigation has the advantage of introducing a predictive component to this methodology. While the network total connectedness measure identifies increases in regional uncertainty associated with major events that shake the markets, the causality relation between total connectedness in different regions – introduced in this paper – provides a quantitative characterization of the flow of uncertainty from region to region, which could be interpreted as the result of contagion. To the best of our knowledge, this causality analysis is the first of its kind.

In future, we will compare this approach with other information theoretic measures with the aim of finding a framework that is capable of qualifying financial uncertainty and its causal effects at all levels of aggregation, from a local single-variable perspective to a global world-market view.

Declaration of interest

The authors report no conflicts of interest. The authors alone are responsible for the content and writing of the paper.

Acknowledgements

T.A. and F.C. acknowledge the support of the UK Economic and Social Research Council (ESRC) in funding the Systemic Risk Centre (ES/K002309/1).

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