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Covariance Matrix Estimation in Time-Varying Factor Models

  • Jingming Yao and Jianhong Wu EMAIL logo
Published/Copyright: July 14, 2025
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Abstract

This paper considers the covariance matrix estimation in time-varying factor models. A so-called two-step method is proposed to estimate the covariance matrix, which is shown to be more accurate for high-dimensional data. Simulation results show that the proposed method has desired performance under finite samples, especially for the case with rapidly changing factor loadings. The empirical study shows that the minimum variance portfolio constructed by this method has desired performance in terms of both return and risk control.

JEL Classification: C13; C23

Corresponding author: Jianhong Wu, Shanghai Normal University, Shanghai 200234, China; and Lab for Educational Big Data and Policymaking, Shanghai 200234, China, E-mail: 

Funding source: National Nature Science Foundation of China

Award Identifier / Grant number: 72173086

Acknowledgments

We are deeply grateful to Professor Jeremy Piger and an anonymous referee for valuable comments that led to substantial improvement of this paper.

  1. Conflict of interest: The authors state no conflict of interest.

  2. Research funding: This research is supported in part by the National Nature Science Foundation of China (Grant No. 72173086).

Appendix

Proof of Theorem 1.

As mentioned in Fan et al. (2024), the rate of convergence for the estimator Σ e is related to the sparsity (quantified by s1) as well as the order of estimators for factor loadings and common factors (i.e., ω(T, N, h)). It is worth noting that this paper uses the Lppca method to estimate the common factors and the time-varying factor loadings. Under Assumptions 13, the order of estimators for factor loadings and common factors are obtained by following from Lemmas C.4 and C.12 of Jung (2021). Further, we obtain ω ( T , N , h ) = 1 N + log N T T h + h 2 log T + J η , which is different from Fan et al. (2024). The solution to 1-penalized log-likelihood estimation problem (6) could also be treated as a particular type of M-estimator for the covariance matrix Σ e , based on minimizing a Bregman divergence between positive definite matrices. Similar to Fan et al. (2024), we have Σ ̂ e Σ e F N + s 1 ω ( T , N , h ) . Conversely, we have Σ ̂ e Σ e Σ ̂ e Σ e = O p d ω ( T , N , h ) , so Σ ̂ e Σ e = O p min { N + s 1 , d } ω ( T , N , h ) . Under Assumption 4 and the assumption that Σ ̂ e Σ e max = O p ω ( T , N , h ) = o p ( 1 ) , the order of precision matrix estimator Σ ̂ e 1 Σ e 1 max = Σ ̂ e 1 Σ e Σ ̂ e Σ e 1 max < Σ e 2 Σ ̂ e Σ e max = O p ω ( T , N , h ) . Similarly, we can derive the rates of convergence for the precision matrix estimator Σ ̂ e 1 under the spectral norm and Frobenius norm, respectively. And then the proof of Theorem 1 is completed.

Proof of Theorem 2.

Following the lead of Fan et al. (2024), we evaluate the norm of the term Σ ̂ y , t Σ y , t in the following way.

For t = 1, 2, …, T, we have the following equalities

Σ y , t = G t Σ F G t + Σ e , Σ ̂ y , t = G ̂ t G ̂ t + Σ ̂ e .

Define a new covariance matrix

Σ ̃ y , t = G t S F G t + Σ e , t = 1,2 , , T ,

where S F = 1 T t = 1 T F ̃ t F ̃ t , and F ̃ t = F t K t , h 1 / 2 .

To estimate the convergence rate of Σ ̂ y , t Σ y , t , we can perform the following scaling

(A.1) sup t Σ ̂ y , t Σ y , t Σ y , t 2 C sup t Σ ̂ y , t Σ ̃ y , t Σ y , t + Σ ̃ y , t Σ y , t Σ y , t .

Here, the norm  Σ y , t 2 is defined as follows: For any N-dimensional matrix A,

(A.2) A Σ y , t 2 = 1 N Σ y , t 1 / 2 A Σ y , t 1 / 2 F 2 C N A F 2 ,

where C is a constant.

For the first term on the right side of (A.1), we decompose it again

(A.3) sup t Σ ̂ y , t Σ ̃ y , t Σ y , t 2 C sup t G ̂ t G ̂ t G t S F G t Σ y , t 2 + Σ ̂ e Σ e Σ y , t 2 .

According to (A.2) and Theorem 1, we can prove that

(A.4) sup t Σ ̂ e Σ e Σ y , t 2 = O p sup Σ ̂ e Σ e F 2 = O p min ( N + s 1 ) , d 2 ω 2 ( T , N , h ) .

Similar to Fan et al. (2024), we have

(A.5) sup t G ̂ t G ̂ t G t S F G t Σ y , t 2 = O p N ω 4 ( T , N , h ) + ω 2 ( T , N , h ) .

It follows from (A.4) and (A.5) that

(A.6) sup t Σ ̂ y , t Σ ̃ y , t Σ y , t 2 = O p N ω 4 ( T , N , h ) + min ( N + s 1 ) , d 2 ω 2 ( T , N , h ) .

For the second term Σ ̃ y , t Σ y , t , following arguments analogous to the proof of Lemma B.10 of Jung (2021), we have

(A.7) sup t Σ ̃ y , t Σ y , t Σ y , t 2 = sup t G t S F Σ F G t Σ y , t 2 = 1 p sup t Σ y , t 1 / 2 G t S F Σ F G t Σ y , t 1 / 2 F 2 1 N sup t Σ y , t 1 / 2 G t F 2 S F Σ F 2 = O p 1 N log ( N T ) T h + h 2 2 = o P N ω 4 ( T , N , h ) + min ( N + s 1 ) , d 2 ω 2 ( T , N , h ) .

Therefore, the final evaluation of Σ ̂ y , t Σ y , t is derived by combining (A.6) and (A.7),

(A.8) sup t Σ ̂ y , t Σ y , t Σ y , t 2 = O p N ω 4 ( T , N , h ) + min ( N + s 1 ) , d 2 ω 2 ( T , N , h ) .

Next, we show the order of Σ ̂ y , t Σ y , t under the norm. We decompose this term in the same way as (A.1)

sup t Σ ̂ y , t Σ y , t max sup t Σ ̂ y , t Σ ̃ y , t max + sup t Σ ̃ y , t Σ y , t max .

The first term on the right side of the inequality is

sup t Σ ̂ y , t Σ ̃ y , t max sup t Σ ̂ e Σ e max + sup t G ̂ t G ̂ t G t Σ F G t max .

By Theorem 1 we have sup t Σ ̂ e Σ e max = O p ω ( T , N , h ) and by the same argument of Wang et al. (2021), we have sup t G ̂ t G ̂ t G t Σ F G t max = O p ω ( T , N , h ) , we know that

(A.9) sup t Σ ̂ y , t Σ ̃ y , t max = O p ω ( T , N , h ) .

Using a similar proof to (A.7), we have

(A.10) sup t Σ ̃ y , t Σ y , t max = O p log ( N T ) T h + h 2 .

So it follows from (A.9) and (A.10) that the convergence rate of Σ ̂ y , t Σ y , t under the norm,

(A.11) sup t Σ ̂ y , t Σ y , t max = O p ω ( T , N , h ) .

Finally, we evaluate the rate of convergence for the precision matrix Σ ̂ y , t 1 . Similarly, we utilize the same decomposition by the triangle inequality,

Σ ̂ y , t 1 Σ y , t 1 Σ ̂ y , t 1 Σ ̃ y , t 1 + Σ ̃ y , t 1 Σ y , t 1 .

By applying the Sherman-Morrison-Woodbury formula to Σ ̃ y , t 1 and Σ y , t 1 , we have

Σ y , t 1 = Σ e 1 Σ e 1 G t Σ F 1 + G t Σ e 1 G t 1 G t Σ e 1 ,
Σ ̃ y , t 1 = Σ e 1 Σ e 1 G t S F 1 + G t Σ e 1 G t 1 G t Σ e 1 .

Then

(A.12) Σ ̃ y , t 1 Σ y , t 1 = Σ e 1 G t Q G t Σ e 1 ,

where Q = S F 1 + G t Σ e 1 G t 1 Σ F 1 + G t Σ e 1 G t 1 Q 1 1 Q 2 1 .

(A.13) Q 1 1 Q 2 1 = Q 1 1 Q 2 Q 1 Q 2 1 Q 1 1 Q 2 1 + Q 2 Q 1 Q 2 Q 1 Q 2 1 .

Then we have

(A.14) Q = Q 1 1 Q 2 1 Q 1 Q 2 Q 2 1 2 1 Q 1 Q 2 Q 2 1 .

By (A.7), we can obtain

(A.15) sup t Q 1 Q 2 = sup t S F 1 Σ F 1 = O p log ( N T ) T h + h 2 .

Then from (A.12) and (A.16), we obtain

(A.16) sup t Σ ̃ y , t 1 Σ y , t 1 = O p 1 N log ( N T ) T h + h 2 = o p min { N + s 1 , d } ω ( T , N , h ) .

Next, we consider Σ ̃ y , t 1 Σ ̂ y , t 1 . By applying the Sherman-Morrison-Woodbury formula, we obtain

Σ ̂ y , t 1 = Σ ̂ e 1 Σ ̂ e 1 G ̂ t I m + G ̂ t Σ ̂ e 1 G ̂ t 1 G ̂ t Σ ̂ e 1 ,
Σ ̃ y , t 1 = Σ e 1 Σ e 1 G t H Σ F H 1 + G t H Σ e 1 G t H 1 G t H Σ e 1 ,

where G t H = G t H t 1 and Σ F H = H t Σ F H t .

Through the same decomposition as in the proof of Theorem 3 in Wang et al. (2021), we have

Σ ̂ y , t 1 Σ ̃ y , t 1 = = 1 6 D ,
D 1 = Σ e 1 Σ ̂ e 1 , D 2 = Σ ̂ e 1 Σ e 1 G ̂ t I m + G ̂ t Σ ̂ e 1 G ̂ t 1 G ̂ t Σ ̂ e 1 , D 3 = Σ e 1 G ̂ t I m + G ̂ Σ ̂ e 1 G ̂ t 1 G ̂ t Σ ̂ e 1 Σ e 1 , D 4 = Σ e 1 G ̂ t G t I m + G ̂ t Σ ̂ e 1 G ̂ t 1 G ̂ t Σ e 1 , D 5 = Σ e 1 G t H I m + G ̂ t Σ ̂ e 1 G ̂ t 1 G ̂ t G t H Σ e 1 , D 6 = Σ e 1 G t H Q ̃ H G t H Σ e 1 ,

with Q ̃ = I m + G ̂ t Σ ̂ e 1 G ̂ t 1 Σ F H 1 + G t H Σ e 1 G t H 1 Q ̃ 1 1 Q ̃ 2 1 .

According to Theorem 1, we have

sup t D 1 = sup t Σ ̂ e 1 Σ e 1 = O p min { N + s 1 , d } ω ( T , N , h ) .

For D2,

sup t D 2 sup t Σ ̂ e 1 Σ e 1 G ̂ t I m + G ̂ t Σ ̂ e 1 G ̂ t 1 G ̂ t Σ ̂ e 1 = O p min { N + s 1 , d } ω ( T , N , h ) .

The last equality uses the fact that sup t I m + G ̂ t Σ ̂ e 1 G ̂ t 1 = O p 1 N .

We can derive the order of sup t D3‖ in a similar way,

sup t D 3 = O p min { N + s 1 , d } ω ( T , N , h ) .

For D4,

sup t D 4 sup t Σ e 1 G ̂ t G t H I m + G ̂ t Σ ̂ e 1 G ̂ t 1 G ̂ t Σ ̂ e 1 = o p ω ( T , N , h ) .

The last equality uses Lemma C.12 in Jung (2021), that is max 1 i N sup t G ̂ t H t 1 G t 2 = o p ω 2 ( T , N , h ) .

The order of sup t D 5 is obtained similarly,

sup t D 5 = O p ω ( T , N , h ) .

For Q ̃ , by using (A.14), we can obtain

sup t Q ̃ = sup t Q ̃ 1 1 Q ̃ 2 1 Q ̃ 1 Q ̃ 2 Q ̃ 2 1 2 1 Q ̃ 1 Q ̃ 2 Q ̃ 2 1 .

From (S15), (S21) of Wang et al. (2021) and Theorem 1, we can conclude that

sup t Q ̃ = O p min { N + s 1 , d } N ω ( T , N , h ) .

Thus,

sup t D 6 = O p min { N + s 1 , d } ω ( T , N , h ) .

In summary, we can get

(A.17) sup t Σ ̂ y , t 1 Σ ̃ y , t 1 = O p min { N + s 1 , d } ω ( T , N , h ) .

Combining (A.8), (A.11) and (A.17), the proof of Theorem 2 is complete.

References

Ao, M., L. Yingying, and X. Zheng. 2019. “Approaching Mean-Variance Efficiency for Large Portfolios.” Review of Financial Studies 32 (7): 2890–919. https://doi.org/10.1093/rfs/hhy105.Search in Google Scholar

Bai, J. 2003. “Inferential Theory for Factor Models of Large Dimensions.” Econometrica 71 (1): 135–71. https://doi.org/10.1111/1468-0262.00392.Search in Google Scholar

Bai, J., and S. Ng. 2002. “Determining the Number of Factors in Approximate Factor Models.” Econometrica 70 (1): 191–221. https://doi.org/10.1111/1468-0262.00273.Search in Google Scholar

Bickel, P. J., and E. Levina. 2008. “Covariance Regularization by Thresholding.” Annals of Statistics 36 (6): 2577–604. https://doi.org/10.1214/08-aos600.Search in Google Scholar

Cai, T., and W. Liu. 2011. “Adaptive Thresholding for Sparse Covariance Matrix Estimation.” Journal of the American Statistical Association 106 (494): 672–84. https://doi.org/10.1198/jasa.2011.tm10560.Search in Google Scholar

Connor, G., M. Hagmann, and O. Linton. 2012. “Efficient Semiparametric Estimation of the Fama–French Model and Extensions.” Econometrica 80 (2): 713–54.10.3982/ECTA7432Search in Google Scholar

Ding, Y., Y. Li, and X. Zheng. 2021. “High Dimensional Minimum Variance Portfolio Estimation Under Statistical Factor Models.” Journal of Econometrics 222 (1): 502–15. https://doi.org/10.1016/j.jeconom.2020.07.013.Search in Google Scholar

Fan, J., Y. Liao, and M. Mincheva. 2011. “High Dimensional Covariance Matrix Estimation in Approximate Factor Models.” Annals of Statistics 39 (6): 3320–56. https://doi.org/10.1214/11-aos944.Search in Google Scholar

Fan, J., Y. Liao, and M. Mincheva. 2013. “Large Covariance Estimation by Thresholding Principal Orthogonal Complements.” Journal of the Royal Statistical Society – Series B: Statistical Methodology 75 (4): 603–80. https://doi.org/10.1111/rssb.12016.Search in Google Scholar PubMed PubMed Central

Fan, J., Y. Liao, and W. Wang. 2016. “Projected Principal Component Analysis in Factor Models.” Annals of Statistics 44 (1): 219–54. https://doi.org/10.1214/15-aos1364.Search in Google Scholar

Fan, Q., R. Wu, Y. Yang, and W. Zhong. 2024. “Time-Varying Minimum Variance Portfolio.” Journal of Econometrics 239 (2): 105339. https://doi.org/10.1016/j.jeconom.2022.08.007.Search in Google Scholar

Jagannathan, R., and T. Ma. 2003. “Risk Reduction in Large Portfolios: Why Imposing the Wrong Constraints Helps.” The Journal of Finance 58 (4): 1651–83. https://doi.org/10.1111/1540-6261.00580.Search in Google Scholar

Johnstone, I. M. 2001. “On the Distribution of the Largest Eigenvalue in Principal Components Analysis.” Annals of Statistics 29 (2): 295–327. https://doi.org/10.1214/aos/1009210544.Search in Google Scholar

Jung, J. 2021. “Essay on Time-Varying Covariance Matrix Estimations with Time-Varying Factor Models.” PhD thesis. Rutgers The State University of New Jersey, School of Graduate Studies.Search in Google Scholar

Lam, C., and J. Fan. 2009. “Sparsistency and Rates of Convergence in Large Covariance Matrix Estimation.” Annals of Statistics 37 (6B): 4254–78. https://doi.org/10.1214/09-aos720.Search in Google Scholar

Ma, S., and L. Su. 2018. “Estimation of Large Dimensional Factor Models with an Unknown Number of Breaks.” Journal of Econometrics 207 (1): 1–29. https://doi.org/10.1016/j.jeconom.2018.06.019.Search in Google Scholar

Markowitz, H. 1952. “Portfolio Selection.” The Journal of Finance 7 (1): 77–91. https://doi.org/10.1111/j.1540-6261.1952.tb01525.x.Search in Google Scholar

Merton, R. C. 1972. “An Analytic Derivation of the Efficient Portfolio Frontier.” Journal of Financial and Quantitative Analysis 7 (4): 1851–72. https://doi.org/10.2307/2329621.Search in Google Scholar

Motta, G., C. M. Hafner, and R. von Sachs. 2011. “Locally Stationary Factor Models: Identification and Nonparametric Estimation.” Econometric Theory 27 (6): 1279–319. https://doi.org/10.1017/s0266466611000053.Search in Google Scholar

Stock, J. H., and M. W. Watson. 2002. “Forecasting Using Principal Components from a Large Number of Predictors.” Journal of the American Statistical Association 97 (460): 1167–79. https://doi.org/10.1198/016214502388618960.Search in Google Scholar

Su, L., and X. Wang. 2017. “On Time-Varying Factor Models: Estimation and Testing.” Journal of Econometrics 198 (1): 84–101. https://doi.org/10.1016/j.jeconom.2016.12.004.Search in Google Scholar

Wang, L., and J. Wu. 2022. “Estimation of High-Dimensional Factor Models with Multiple Structural Changes.” Economic Modelling 108: 105743. https://doi.org/10.1016/j.econmod.2021.105743.Search in Google Scholar

Wang, H., B. Peng, D. Li, and C. Leng. 2021. “Nonparametric Estimation of Large Covariance Matrices with Conditional Sparsity.” Journal of Econometrics 223 (1): 53–72. https://doi.org/10.1016/j.jeconom.2020.09.002.Search in Google Scholar


Supplementary Material

This article contains supplementary material (https://doi.org/10.1515/snde-2025-0042).


Received: 2025-03-27
Accepted: 2025-06-30
Published Online: 2025-07-14

© 2025 Walter de Gruyter GmbH, Berlin/Boston

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