The Endowment Model and Modern Portfolio Theory

We develop a dynamic portfolio-choice model with illiquid alternative assets to analyze the “endowment model,” widely adopted by institutional investors such as pension funds, university endowments, and sovereign wealth funds. In the model, the alternative asset has a lock-up, but can be liquidated at any time by paying a proportional cost. We model how investors can engage in liquidity diversification by investing in multiple illiquid alternative assets with staggered lock-up expirations, and show that doing so increases alternatives allocations and investor welfare. We show how illiquidity from lock-ups interacts with illiquidity from secondary market transaction costs resulting in endogenous and time-varying rebalancing boundaries. We extend the model to allow crisis states and show that increased illiquidity during crises causes holdings to deviate significantly from target allocations.

For helpful comments, we thank an anonymous Associate Editor, two anonymous referees, Patrick Bolton, Bruno Biais (Editor), Winston Dou, Thomas Gilbert, Harrison Hong, Steve Kaplan, Monika Piazzesi, Jim Poterba, Tom Sargent, Mark Schroder, and Luis Viceira; seminar participants at Columbia University; and participants at the European Finance Association and NBER New Developments in Long-Term Asset Management conferences. We thank Matt Hamill and Ken Redd of NACUBO and John Griswold and Bill Jarvis of Commonfund for assistance with data. Dimmock gratefully acknowledges financial support from the Singapore Ministry of Education research grant R-315-000-133-133. The views expressed herein are those of the authors and do not necessarily reflect the views of the National Bureau of Economic Research.

MARC RIS BibTeΧ

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  • February 13, 2019
  • June 24, 2021
  • December 29, 2021

Published Versions

Stephen G. Dimmock & Neng Wang & Jinqiang Yang, 2024. " The Endowment Model and Modern Portfolio Theory, " Management Science, vol 70(3), pages 1554-1579.

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Portfolio Theory and Management

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Portfolio Theory and Management

2 Modern Portfolio Theory

  • Published: February 2013
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This chapter surveys modern portfolio theory, which is one of the most spectacular developments of finance in the last 50 years. It starts with the basic one-period setup under the assumption of normality with the successive contributions including the basic Markowitz mean-variance framework, the efficient frontier, and the Sharpe-Lintner capital asset pricing model. Utility and risk aversion are also discussed. The chapter then discusses the multiperiod extension and Merton's optimal asset allocation. The second part of the chapter shows how to extend the framework to allow for parameter uncertainty. In that process, the chapter also briefly reviews needed concepts such as the predictive density, shrinkage, and how the Bayesian framework allows the incorporation of prior views to improve on the precision of estimates necessary in the portfolio construction process.

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Model-based vs. agnostic methods for the prediction of time-varying covariance matrices

  • Original Research
  • Published: 13 September 2024

Cite this article

portfolio theory research paper

  • Jean-David Fermanian 1   na1 ,
  • Benjamin Poignard 2 , 3   na1 &
  • Panos Xidonas   ORCID: orcid.org/0000-0003-3325-1474 4   na1  

This article is written in memory of Harry Markowitz, the founder of modern portfolio theory. We report a few human perspectives of his character, we review a large number of his contributions, published both in operations research and finance oriented journals, and we focus on one of the most critical, and still open, portfolio theory issues, the forecast of covariance matrices. Our contribution in this paper is placed exactly towards this direction. More specifically, we compare the performances of several approaches to predict the variance-covariance matrices of vectors of asset returns, through simulated and real data experiments: some dynamic models such as Dynamic Conditional Correlation (DCC) and C-vine GARCH on one side, and several agnostic methods (Average Oracle, usual “Sample” matrix) on the other side. The most robust methods seem to be DCC and the Average Oracle approaches.

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Jean-David Fermanian, Benjamin Poignard and Panos Xidonas have contributed equally to this work.

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ENSAE-CREST, Finance Department, 5 av. Henry le Chatelier, 91120, Palaiseau, France

Jean-David Fermanian

Osaka University, Graduate School of Economics, 1-7, Machikaneyama, Toyonaka, 560-0043, Osaka, Japan

Benjamin Poignard

RIKEN-AIP, Tokyo, Japan

ESSCA School of Management, 55 quai Alphonse Le Gallo, 92513, Paris, France

Panos Xidonas

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A Vines and partial correlations

This section emphasizes how to specify a relevant set of partial correlations by considering a graphical approach based on vines. Even if it largely takes up and updates Appendix B in Poignard and Fermanian ( 2019 ), we have recalled all these elements for the sake of self-consistency and to help readers. A general presentation of vine models can be found in Czado ( 2019 ).

1.1 A.1 Vines

Let \({{\mathcal {N}}}\) be a set of n elements. By definition, \(T = ({{\mathcal {N}}},{{\mathcal {E}}})\) is a tree with nodes \({{\mathcal {N}}}\) and edges \({{\mathcal {E}}}\) if \({{\mathcal {E}}}\) is a subset of unordered pairs of \({{\mathcal {N}}}\) with no cycle and if there is a path between each pair of nodes. Moreover, vines on n elements are undirected graphs that nest sets of some connected trees \(T_1,\ldots ,T_{n-1}\) , where the edges of tree \(T_j\) are the nodes of tree \(T_{j+1}\) , \(j=1,\ldots ,n-2\) . A regular vine (R-vine) on n elements is a vine in which two edges in tree \(T_j\) are joined by an edge in tree \(T_{j+1}\) only if these edges share a common node, for any \(j=1,\ldots ,n-2\) . A formal definition is given below. See Joe and Kurowicka ( 2011 ) for a survey and additional results.

Definition 1

\(V\left( n\right) \) is a labeled regular vine on n elements if:

\(V\left( n\right) = \left( T_1,T_2,\ldots ,T_{n-1}\right) \) .

\(T_1\) is a connected tree with nodes \({{\mathcal {N}}}_1 = 1,2,\ldots ,n\) and edges \({{\mathcal {E}}}_1\) . For \(i=2,\ldots ,n-1\) , \(T_i\) is a connected tree with nodes \({{\mathcal {N}}}_i = {{\mathcal {E}}}_{i-1}\) , and the cardinality of \({{\mathcal {N}}}_i\) is \(n-i+1\) .

If a and b are nodes of \(T_i\) connected by an edge in \(T_i\) , where \(a = \{a_1,a_2\}\) and \(b = \{b_1,b_2\}\) , then exactly one of the \(a_i\) equals one of the \(b_i\) . This is the proximity condition.

We consider only regular vines in this paper, and the properties we state hereafter are true for such vines implicitly. There are \(n(n-1)/2\) edges in a regular vine on n variables. An edge in tree \(T_j\) is an unordered pair of nodes of \(T_j\) , or equivalently, an unordered pair of edges of \(T_{j-1}\) . The degree of a node is the number of edges incident with it.

Two particular cases of R-vines are important, traditionally. A regular vine is called a canonical vine (C-vine) if each tree \(T_i\) has a unique node of degree \(n-i\) , i.e., a node with maximum degree. A regular vine is called a D-vine if all nodes in \(T_1\) have degree not higher than 2.

The variables reachable from a given edge via the membership relation are called the constraint set of that edge. When two edges are joined by an edge of the next tree, the intersection of the respective constraint sets are the conditioning variables , and the symmetric differences of the constraint sets are the conditioned variables . With the notations of point 3 of the previous definition, at tree \(T_i\) , say \(a_1 = b_1\) , and \(a_1\) is a common element of a and b . This means that, at tree \(T_{i+1}\) , \(a_1\) enters the conditioning set of \(\left( a_2,b_2\right) \) . Thus, we define the conditioning and conditioned sets formally as follows.

Definition 2

For \(e \in {{\mathcal {E}}}_i,\,i \le n-1\) , the constraint set associated with e is the complete union of the elements in \(\{1,\ldots ,n\}\) that are reachable from e by the membership relation. It is denoted by \(U^{\star }_e\) .

Definition 3

For \(i=1,\ldots ,n-1\) , if \(e \in {{\mathcal {E}}}_i\) , it connects two elements j and k in \({{\mathcal {N}}}_i\) and it can be written \(e=\left\{ j, k\right\} \) . The conditioning set associated with e is \(L_e:= U^{\star }_j \cap U^{\star }_k\) , and the conditioned set associated with e is a pair \(\left\{ C_{e,j},C_{e,k}\right\} := \left\{ U^{\star }_j {\setminus } L_e,U^{\star }_k {\setminus } L_e\right\} \) .

Obviously, since the edges of a given tree \(T_i\) are the nodes of \(T_{i+1}\) , the same concepts of constraint/conditioning/conditioned sets apply to all the nodes in a vine.

(Bedford & Cooke, 2002 ) Let a regular vine on n variables. Then,

The total number of edges is \(n(n-1)/2\) ;

Two different edges have different constraint sets;

Each conditioned set is a doubleton and each pair of variables occurs exactly once as a conditioned set;

If \(e\in {{\mathcal {E}}}_i\) , then \(\# U^{\star }_e=i+1\) , \(\# L_e = i-1\) ;

If two edges have the same conditioning set, then they are the same edge.

In a regular vine, the edges of \(T_{m+1}\) (equivalently the nodes of \(T_{m+2}\) ) will be denoted by \(e=(a_j, a_k | b_1,\ldots ,b_m)\) , where \(a_j\) , \(a_k\) and the \(b_l\) , \(l=1,\ldots ,m\) are different elements in \(\{1,\ldots ,n\}\) . This notation means that the conditioning set of e is \(L_e=\{b_1,\ldots ,b_m\}\) , and the conditioned set of e is \(\{a_j,a_k\}\) . Both C-, D- and R-vine and the concepts above can be visualized on Figs.  1 ,  2 and  3 .

To have the intuition, keep in mind that a node represents a random variable, and an edge between two nodes means we will specify the dependence between these two particular nodes, in general through a copula (that will be reduced to a partial correlation hereafter). Such copulae have to be defined afterwards, but, for the moment, assume this can be done easily. Typically, the goal is to describe the joint law of the n asset returns. For instance, in Fig.  1 , the five nodes in \(T_1\) may be the asset returns \(r_{i}\) , \(i=1,\ldots ,5\) , associated to stock indices. The first tree tells us we will specify the dependencies between \(r_1\) and the other returns \(r_i\) , \(i>1\) . Here, we select 1 as the core index (the “main factor”) in this portfolio. Once we have controlled the \(T_1-\) related dependencies, the new nodes in \(T_2\) are conditional asset returns given \(r_1\) . We select asset 2 given 1 as the “most relevant” one. The new edges tell us we focus now on conditional copulae between the latter node and the returns \(r_j\) given \(r_1\) , \(j=2,\ldots ,5\) . And we go on with \(T_3\) , dealing with the asset returns \(r_j\) given \(r_1\) and \(r_2\) , \(j=3,4,5\) , etc. With such a C-vine and a set of convenient bivariate copulae, we obtain the joint law of \((r_1,\ldots ,r_5)\) by gathering and multiplying conveniently all the (conditional) copulae we haver considered above. This is the simplest way of building vines. Obviously, more complex structures may be relevant too, as in the R-vine of Fig.  3 . With heterogeneous portfolios, for instance, it would be fruitful to particularize several nodes in \(T_1\) . See Aas et al. ( 2009 ) for other insights. In terms of model specification, the first chosen trees are crucial because they correspond to our intuitions (our “priors”) about the most important linkages among the assets in the portfolio. Moreover, from some level on and in practice, it is often possible and useful to assume no dependencies: see the “r-vine free” property in Definition  5 below.

figure 1

Example of a C-vine on five variables. Lecture: the two nodes (1, 2) and (1, 3) in \(T_2\) are connected by the edge (2, 3|1), whose constraint set is \(\{1,2,3\}\) , conditioned set is \(\{2,3\}\) and conditioning set is \(\{1\}\)

figure 2

Example of a D-vine on five variables. Lecture: the two nodes (1, 3|2) and (2, 4|3) in \(T_3\) are connected by the edge (1, 4|2, 3), whose constraint set is \(\{1,2,3,4\}\) , conditioned set is \(\{1,4\}\) and conditioning set is \(\{2,3\}\)

figure 3

Example of a R-vine on five variables. The solid, dotted, dashed-dotted and black solid lines correspond to the edges of \(T_1\) , \(T_2\) , \(T_3\) and \(T_4\) respectively

The next section focuses on how such vines are related to some subsets of the partial correlations that are associated to a random vector.

1.2 A.2 Partial correlations

Let \({{\varvec{X}}}=(X_1,\ldots ,X_n)\) be a n -dimensional random vector, \(n\ge 2\) , with zero mean. For any indices i ,  j in \( \{1,\ldots ,n\}\) , \(i\ne j\) and any subset \(L\subset \{1,\ldots ,n\}\) , for which i and j do not belong to L , \(\rho _{i,j | L}\) is called the partial correlation of \(X_i\) and \(X_j\) , given \(X_k\) , \(k\in L\) . It is the correlation between the orthogonal projections of \(X_i\) and \(X_{j}\) on \(< X_k, k\in L >^{\perp }\) , the orthogonal of the subspace generated by \(\{X_k,\,k\in L\}\) . When L is empty, then \(\rho _{i,j|\emptyset } = \rho \left( X_i,X_j\right) :=\rho _{i,j}\) is the usual correlation. Note that, if the random vector \({{\varvec{X}}}\) is normal, then its partial correlations correspond to some conditional correlations.

Interestingly, partial correlations can be computed from usual correlations with a recursive formula. Let \(\left( i,j,k\right) \) be any set of distinct indices, and L be another (possibly empty) set of indices that is disjoint from \(\left( i,j,k\right) \) . Following Lewandowski et al. ( 2009 ), we have

Assume we know the usual correlations \(\rho _{i,j}\) , for any couple ( i ,  j ), \(i\ne j\) . We check easily that any partial correlation can be calculated by invoking ( .1 ) several times with increasing subsets L . Actually, the opposite property is true if we start from a convenient subset of partial correlations. Indeed, the edges of a regular vine on n elements may be associated with the partial correlations of a n -dimensional random vector in the following way: for \(i = 1,\ldots ,n-1\) , consider any \(e \in {{\mathcal {E}}}_i\) , the set of edges at tree \(T_i\) . Let \(\left\{ j,k\right\} \) be the two conditioned variables of e , and \(L_e\) its conditioning set. We associate the partial correlation \(\rho _{j,k|L_e}\) to this node. Kurowicka and Cooke ( 2006 ) call this structure a partial correlation vine specification , that is simply a R-vine for which any edge is associated to a number in \(\left]-1,1\right[\) . Actually, all positive definite correlation matrices may be generated by setting a (fixed) R-vine on n variables, and by assigning different partial correlations to all the nodes of this vine. This means setting \(\rho _{e}\) to any \(e \in \overset{n-1}{\underset{i=1}{\cup }} {{\mathcal {E}}}_i \) , and these partial correlations may be chosen in \(\left]-1,1\right[\) arbitrarily . This is the content of Corollary 7.5 in Bedford and Cooke ( 2002 ).

(Bedford & Cooke, 2002 ) For any regular vine on n elements, there is a one-to-one mapping between the set of \(n \times n\) positive definite correlation matrices and the set of partial correlation specifications for the vine.

In other words, any set of \(n(n-1)/2\) partial correlations that are deduced from a regular vine induce a true correlation matrix. Actually, the formulas ( .1 ) above enable to build such \(n\times n\) correlation matrices based on \(n(n-1)/2\) arbitrarily chosen partial correlations : see Kurowicka and Cooke ( 2003 ), Joe ( 2006 ). For a given partial correlation vine, some explicit algorithms map the (usual) correlations and the underlying partial correlations: see Lewandowski et al. ( 2009 ). Such algorithms are available in the R-package called “vine-copula”. See Brechmann and Schepsmeier ( 2013 ), for instance.

Definition 4

Let a vine \(V(n)=(T_1,T_2,\ldots ,T_{n-1})\) . The set of partial correlations associated to this vine is denoted by \({\tilde{C}}_{V(n)}:=\left( C(T_1),C(T_2),\ldots ,C(T_{n-1})\right) \) . Denote by \(R\left( {\tilde{C}}_{V(n)}\right) \) the set of usual correlations that are deduced from \({\tilde{C}}_{V(n)}\) .

Theorem  2 means that, whatever the values of the partial correlations \({\tilde{C}}_{V(n)} \) associated to a regular vine V ( n ), we get a true correlation matrix with the coefficients \(R\left( {\tilde{C}}_{V(n)}\right) \) . Since a standardized gaussian random vector is fully specified by its correlation matrix, we obtain its joint law once we have chosen a partial correlation vine specification. At the opposite, for any gaussian vector, there are many corresponding partial correlation vine specifications. In a gaussian world, we recover the interpretation of vines as descriptors of random vector distributions. But more generally, partial correlation vine specifications can be associated to any random vector, just to describe its correlation matrix (when it exists).

To illustrate these ideas, let us revisit Fig.  1 under a partial correlation point of view: an associated partial correlation vine will specify the set of partial correlations \(\left\{ \rho _{12}, \rho _{13},\rho _{14},\rho _{15}, \rho _{23|1},\rho _{24|1}, \rho _{25|1},\rho _{34|12}, \rho _{35|12},\rho _{45|123}\right\} ,\) that is sufficient to recover the correlation matrix between the five assets. To interpret such numbers, consider linear regressions of some conditioned sets on their conditioning sets. For instance, the node (1, 2) and the node (1, 3) are connected, and the model will specify the partial correlation \(\rho _{12|3}\) . This is the correlation between the residuals of the linear regressions of \(r_2\) and \(r_3\) on \(r_1\) . Roughly, this measures to what extent \(r_2\) and \(r_3\) are “dependent” given \(r_1\) . In practical terms, an econometrician could classify the portfolio components by their (a priori) order of importance. This order may depend on the final phenomenon that is modelled. For instance, if the portfolio payoff depends strongly on emerging markets, it may be relevant to select “Russia” or “Brazil” first instead of “the USA”. Intuitively, the latter strategy is intermediate between a factor model where we would regress any asset return on a few pre-specified ones, and a PCA where the factors are linear combinations of all returns.

This way of interpreting C-vines has to be revisited with D-vines or even general R-vines. Roughly, D-vines are based on an ordered vision of dependencies across asset returns: any asset is associated to one or two neighbors, with whom correlations are relatively strong. Once they are controlled, the main remaining risk is measured by the correlation with (one or) two other known assets, etc. Such a linear view of the strength of dependencies is probably unrealistic in finance. At the opposite, R-vines allow very general and flexible hierarchies and orders among the sequences of partial correlations of interest. Virtually, they allow to integrate any a priori “prior” information, as long as it is consistent with the proximity condition.

For the sake of parsimony, it would be interesting to cancel (or to leave constant, at least) all partial correlations associated to a vine, after some given level r . When zero partial correlations are assumed after the latter level, we would like to know whether the corresponding (usual) correlations depend on the trees \(T_r,T_{r+1},\ldots ,T_{n-1}\) that could be built above.

Definition 5

We say that a vine is r -VF (VF for vine-Free) if

for any alternative vine \(V'(n):= (T_1,T_2,\ldots ,T_{r-1},T'_r,\ldots ,T'_{n-1})\) , where the partial correlations associated to the edges of \(T'_k\) , \(k\ge r\) , are zero.

If a vine is r -VF, once the partial correlations are zero above the level r , the correlations are independent on the way this vine has been built from this level. This r -VF property actually holds for any R-vine. This is a consequence of Theorem 2.3 in Brechmann and Joe ( 2015 ). They observed that the density of an underlying Gaussian vector is not altered when choosing arbitrary trees \(T_{r+1},\ldots ,T_{n-1}\) with associated zero partial correlations.

B Choice of the vine structure and truncation

In this section, we discuss the selection of a suitable vine structure in the case of high-dimensional portfolios (hundreds of assets). To shrink the number of parameters and the numerical noises due to the accumulation of statistical errors with the sequential inference procedure, two approaches can be employed:

The top-down approach of Lewandowski et al. ( 2009 );

The bottom-up approach of Dissmann et al. ( 2013 ).

In case (ii), an algorithm is available in the R vine-copula package. Footnote 3 Nonetheless, its objective is to select the whole model, i.e., the vine structure and the families of bivariate copulas together under the simplifying assumption (Derumigny & Fermanian, 2017 ). This is probably excessive because we simply aim to extract a vine structure that would allow “truncation” after some stage, i.e., to assume bivariate independence copulas in the copula-vine model from some tree on.

Case (i) may be interesting in our context because the truncation order is based on partial correlations and we do not need to specify the bivariate copula families of the model. Nonetheless, the code of this algorithm is currently not available on the web.

Beside, for the arbitrary R-vine with truncation, the approach in Kurowicka ( 2010 ) relies on Pearson product moment correlations. Czado et al. ( 2012 ) proposed an intensive application of some model selection test (Clarke’s test) to choose the best vine structure. In the same spirit, Brechmann et al. ( 2012 ) considered a Vuong test-based procedure, a technique that requires the likelihood function at each tree level.

We could propose an alternative method that should be relevant for C-vine model selection: in a bottom-up approach, select the threshold level based on empirical partial correlations, without depending on any parametric copula family. Indeed, on any node of the vine, say ( i ,  j | D ), it is always possible to empirically evaluate the partial correlation between \(X_i\) and \(X_j\) given the other returns \(X_k\) , \(k\in D\) (calculate the empirical correlation between the residuals of the linear regressions of \(X_i\) or \(X_j\) on \({{\varvec{X}}}_D\) ). This may be questionable because low partial correlations do not mean low levels of dependence strictly speaking. Nonetheless, this method would yield a convenient “proxy” in the case of vine-GARCH models, for which the focus is on partial correlation and not on the joint law of returns. In practice, we could consider that all the bivariate copulas in the vine will be the independence copula from level k when the average partial correlation calculated on all nodes of tree k are lower than a threshold (0.1, for instance).

Finally, it is worth summarizing the three methods implemented in the vine-GARCH package Footnote 4 to select the central nodes for C-vines up to a user-specified level:

The average empirical Kendall’s tau (AKT) between \((z_{kt})_{t=1,\ldots ,T}\) and the other components \((z_{jt})_{t=1,\ldots ,T}\) , \(j\ne k\) , that is \(\max _k\sum _{j \le N, j \ne k}|{{\widehat{\tau }}}_{kj}|\) with \({{\widehat{\tau }}}_{kj}\) the usual empirical estimator of the Kendall’s tau. The variable with the highest AKT is selected as the central node of \(T_1\) ; the central node of \(T_2\) is the variable with the second highest AKT; all other subsequent central nodes are set according to this criterion.

The average linear correlation coefficient (ALC) between \((z_{kt})_{t=1,\ldots ,T}\) and the other components \((z_{jt})_{t=1,\ldots ,T}\) , \(j\ne k\) , that is \(\max _k\sum _{j \le N, j \ne k}|{{\widehat{\rho }}}_{kj}|\) , where \(\rho _{kj}\) denotes the linear correlation coefficient between \(z_{kt}\) and \(z_{jt}\) . Then, the variable with the highest ALC is selected as the central node of \(T_1\) ; the central node of \(T_2\) is the variable with the second highest AKT; etc.

The average conditional Kendall’s tau non-parametric estimator that builds upon the work of Gijbels et al. ( 2015 ), Section 3.2. More precisely, the selection of the central node in tree \(T_1\) is performed by computing \(\max _k\sum _{j \le N, j \ne k}|{{\widehat{\tau }}}_{kj}|\) , where \({{\widehat{\tau }}}_{kj}\) is the empirical estimator of Kendall’s tau. Then, conditional to the variable selected as the central node in tree \(T_1\) , say l , we compute \(\max _{k \in \{1,\ldots ,N\}{\setminus } \{l\}}\sum _{j \in \{1,\ldots ,N\}{\setminus } \{l\}, j \ne k}|{{\widehat{\tau }}}_{kj|l}|\) , where \({{\widehat{\tau }}}_{kj|l}\) is the estimator of the conditional Kendall’s tau computed according to formulas (3.4)-(3.5) in Gijbels et al. ( 2015 ). Then, one can proceed in an iterative manner to select the central nodes in \(T_3,\ldots ,T_{N-1}\) , where the conditional Kendall’s tau will be \({{\widehat{\tau }}}_{kj|L}\) with \(|L|>1\) . When one aims to select all of the central nodes of the vine, the method may actually be time-consuming and unstable when the dimension N is large since the conditioning set of indices L in \({{\widehat{\tau }}}_{kj|L}\) becomes larger, which alters the precision of the non-parametric estimator (indeed, it involves some kernel smoothing products that may be unstable for L large). Therefore, this method should be used with truncation, that is when it is reasonable to assume some level r , typically \(r=3,4\) , from where the structure of the C-vine can be arbitrarily set as it does not alter the computation of the classic correlations due to the r -vine free property. The choice of r is clearly user-specified and depends on the data.

C Practical considerations

In light of the curse of dimensionality problem, we propose to rely on the iterative node-by-node procedure described in Section 3.3 in Poignard and Fermanian ( 2019 ) for the estimation of the simplified C-vine GARCH model: 3 parameters only need to be estimated for each partial correlation process; a recursion is employed up to the last tree in the non-truncated case or up to a user-specified tree (usually up to \(T_2\) or \(T_3\) ) in the truncated case. In the non-truncated case, this iterative procedure consists of \(N(N-1)/2\) optimization problems, corresponding to the estimation of the bivariate dynamics associated with any edge of the vine: the \(N-1\) edges of the first tree (and, thus, the correlation dynamics of \(T_1\) ) can be estimated independently. Then, the \(N-2\) partial correlation dynamics of \(T_2\) can be estimated independently, given the dynamic partial correlation processes estimated in \(T_1\) . In the truncated case, this iterative procedure is performed up to the desired level of truncation, i.e., the level from which the edges are set as zero or as constant (mean) partial correlations. This estimation strategy allows to break the curse of dimensionality. Indeed, the estimation of a non-truncated simplified C-vine model necessitates the “brute-force” full estimation of \(O(N^2)\) parameters by optimization, a typically unfeasible task in practice. Note that the node-by-node procedure can be employed in the simplified C-vine model only, due to the absence of cross-effects.

In Table  19 , we report the time required for the estimation of each variance-covariance model and the average oracle for S&P 500 and MSCI portfolios. For both datasets, we use the same in-sample periods as in Sect.  4 . These figures were obtained on a Mac-OS Apple M1 Ultra with 20 cores and 128 GB Memory. The version of the Matlab software is 9.12.0.1975300 (R2022a) Update 3 equipped with the Parallel Computing Toolbox, Version 7.6. To assess the sensitivity of the iterative procedure for the C-vine GARCH, the estimation is performed up to trees \(T_3\) and \(T_4\) with the following structure:

MSCI: when up to \(T_3\) , we set USA in \(T_1\) , Germany in \(T_2\) and Japan in \(T_3\) ; when up to \(T_4\) , we set USA in \(T_1\) , Germany in \(T_2\) , Japan in \(T_3\) and Italy in \(T_4\) .

S&P 500: when up to \(T_3\) , we set Berkshire Hathaway in \(T_1\) , JPMorgan in \(T_2\) and Apple in \(T_3\) ; when up to \(T_4\) , we set Berkshire Hathaway in \(T_1\) , JPMorgan in \(T_2\) , Apple in \(T_3\) and Exxon in \(T_4\) .

The time requirement for estimating the parametric C-vine GARCH is larger due to the need to compute, for each t , the innovation variables of the partial correlation process. The latter task requires the conditional marginal variances and the conditional correlations, which are deduced from partial correlations. This stage is not required in the non-parametric approach. The figures in Table  19 relate to time estimation only, not the time required for generating the correlation matrix from the partial correlation matrix process, which is independent of the truncation level. For the MSCI portfolio, this time approximately 5 s, whereas for the S&P 500 portfolio, it is 19 min and 30 s.

Concerning AO, the computation cost comes from the diagonalizations of B empirical ( N ,  N )-covariance matrices. Using an usual method as the QR algorithm or the “Divide and Conquer” method, the cost a obtaining the eigenvalues of a single matrix is \(O(N^3)\) (Pan & Chen, 1999 ). Thus, taking into account the estimation stage of these B covariance matrices, implementing the AO method has a cost of order \(O(BN^2T+BN^3)\) .

Another point worth mentioning is the treatment of missing values in financial datasets, even if we have not met missing values in Sect.  4 . This problem has fuelled a large amount of academic literature: see the reference textbooks Graham ( 2012 ), Little and Rubin ( 2019 ), for instance. In the case of covariance matrix estimation, pairwise deletion can be applied (Kim & Curry, 1977 ), followed by some projection on the space of positive definite matrices. Beside, numerous “imputation” of “completion” methods have been proposed to fill the gaps directly in datasets: "last observation carried forward", linear or Brownian bridge interpolation, nearest neighbors averaging, singular spectrum analysis, Iterative PCA, etc. See dos Santos ( 2021 ) and the references therein, e.g. Notably, Cao et al. ( 2018 ) proposed to impute missing values with bidirectional recurrent dynamics obtained through some neural networks. Missing values can also been predicted by fitting a parametric data-generating model with the observations, at the price of misspecification risk: Amelia (Honaker et al., 2011 ), multivariate imputation by chained equations (Van Buuren & Oudshoorn, 1999 ), etc. See the survey of Fung ( 2006 ).

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Fermanian, JD., Poignard, B. & Xidonas, P. Model-based vs. agnostic methods for the prediction of time-varying covariance matrices. Ann Oper Res (2024). https://doi.org/10.1007/s10479-024-06238-4

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    Portfolio theory is a well-developed paradigm. There are excellent textbooks on the subject. Of course, we are especially partial to our own Modern Portfolio Theory and Investment Analysis. There are also good reviews in more advanced doctoral-level texts such as Ingersoll (1987) or Huang and Litzenberger (1988).

  17. Modern Portfolio Theory: A Review of the Work Done on Performance

    This system called the "Modern Portfolio Theory" which emerged in 1960‟s is considered in this review. Even though the theory faced a lot of criticism, it still remains the principal foundation for the financial theory. The aim of this review is to examine whether Modern Portfolio Theory is an effective tool for portfolio management. II.

  18. Model-based vs. agnostic methods for the prediction of time-varying

    This article is written in memory of Harry Markowitz, the founder of modern portfolio theory. We report a few human perspectives of his character, we review a large number of his contributions, published both in operations research and finance oriented journals, and we focus on one of the most critical, and still open, portfolio theory issues, the forecast of covariance matrices. Our ...

  19. PDF The modern portfolio theory as an investment decision tool

    This research paper is academic exposition into the modern portfolio theory (MPT) written with a primary objective of showing how it aids an investor to classify, estimate, and control both the kind and the amount of expected risk and return in an attempt to maximize portfolio expected return for a given

  20. PDF FOUNDATIONS OF PORTFOLIO THEORY

    FOUNDATIONS OF PORTFOLIO THEORY. Nobel Lecture, December 7, 1990. by. HARRY M. MARKOWITZ. Baruch College, The City University of New York, New York, USA. When I studied microeconomics forty years ago, I was first taught how optimizing firms and consumers would behave, and then taught the nature of the economic equilibrium which would result ...

  21. (PDF) The Markowitz Portfolio Theory

    Chief Editor. Download Free PDF. View PDF. Portfolio Theory: The Contribution of Markowitz's Theory to Information System Area. Antônio Carlos G Maçada, Pietro Dolci. Portfolio theory is concerned with risk and return. However, assigning weight to the risk at least equal to the yield was the big news in the 1950s.

  22. Understanding Modern Portfolio Construction by Cullen O. Roche

    Abstract. Over the last 75 years there have been great strides in modern finance, portfolio theory and asset allocation strategies. Despite this progress the process of portfolio construction remains grounded in many theoretical concepts that can result in inappropriate or unrealistic frameworks.

  23. PDF Portfolio Selection Harry Markowitz The Journal of Finance, Vol ...

    * This paper is based on work done by the author while at the Cowles Commission for Research in Economics and with the financial assistance of the Social Science Research Council. It will be reprinted as Cowles Commission Paper, New Series, No. 60. 1. See, for example, J.B. Williams, The Theory of Investment Value (Cambridge, Mass.:

  24. Portfolios for long-term investors

    CONTACT. Portfolios for long-term investors. Research. Jan 13. Written By John Cochrane. Jan 13 2022. Review of Finance,26(1), 1-42 (2022). This is an essay on portfolio theory and practice, which evolved from a keynote talk at the NBER conference, ``New Developments in Long-Term Asset Management" Jan 21 2021.

  25. Prospective teachers' views and experiences with e-portfolios

    Student and teacher perceptions of e-portfolios. Research on perceptions of e-portfolios typically either focuses on the learner or teacher. For example, an online quantitative study by Ciesielkiewicz (Citation 2019) was undertaken with the goal of gathering 121 ITE students' perspectives on the value and usefulness of e-portfolios.As an important piece of research, this study found that the ...