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Inferring bi-directional interactions between circadian clock genes and metabolism with model ensembles

  • Marco Grzegorczyk ORCID logo EMAIL logo , Andrej Aderhold und Dirk Husmeier
Veröffentlicht/Copyright: 14. Februar 2015
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Statistical Applications in Genetics and Molecular Biology
Aus der Zeitschrift Statistical Applications in Genetics and Molecular Biology Band 14 Heft 2

Abstract

There has been much interest in reconstructing bi-directional regulatory networks linking the circadian clock to metabolism in plants. A variety of reverse engineering methods from machine learning and computational statistics have been proposed and evaluated. The emphasis of the present paper is on combining models in a model ensemble to boost the network reconstruction accuracy, and to explore various model combination strategies to maximize the improvement. Our results demonstrate that a rich ensemble of predictors outperforms the best individual model, even if the ensemble includes poor predictors with inferior individual reconstruction accuracy. For our application to metabolomic and transcriptomic time series from various mutagenesis plants grown in different light-dark cycles we also show how to determine the optimal time lag between interactions, and we identify significant interactions with a randomization test. Our study predicts new statistically significant interactions between circadian clock genes and metabolites in Arabidopsis thaliana, and thus provides independent statistical evidence that the regulation of metabolism by the circadian clock is not uni-directional, but that there is a statistically significant feedback mechanism aiming from metabolism back to the circadian clock.

Keywords: circadian clock; determination of time delays; model ensembles; plant metabolism; regulatory network inference
Corresponding author: Marco Grzegorczyk, Johann Bernoulli Institute (JBI), Faculteit Wiskunde en Natuurwetenschappen (FWN), Groningen University, Nijenborgh 9, Postbus 407, 9747 AG Groningen 9700 AK, The Netherlands, e-mail:

Acknowledgments

The work described in the present article is part of the TiMet project on linking the circadian clock to metabolism in plants. TiMet is a collaborative project (Grant Agreement 245143) funded by the European Commission FP7, in response to call FP7-KBBE-2009-3. A.A. is supported by the TiMet project. Parts of the work were done while M.G. was supported by the German Research Foundation (DFG), research grant GR3853/1-1. We would like to thank Catherine Higham for helpful discussions.

Appendix

Marginal likelihood for time delays

In the hierarchical Bayesian regression (HBR) model, described in Subsection 2.5 of Aderhold et al. (2014), the target observations yg are modelled independently for each target g (g=1, …, G):

(13)yg|(Xπg, σg2, wg, πg)~N(XπgTwg, σg2I) (13)

where wg are the regression parameter vectors, determined by the sets of regulators (covariates) πg, Xπg is the regressor matrix implied by the regulator set πg, and σg2 is the node-specific noise variance parameter. Gaussian priors are imposed on the regression parameter vectors:

(14)wg|(σg2, δg, πg)~N(0, δgσg2I) (14)

where δg is the target-specific SNR-hyperparameter. The noise variance parameters σg2 and the SNR-hyperparameters δg are assumed to be inverse Gamma distributed with fixed (hyper-)hyperparameters (g=1, …, G). A more detailed model description can be found in Subsection 2.5 of Aderhold et al. (2014). Here, we introduce a new “time lag” parameter τ∈{0, …, τMAX}, which indicates the time lag between the target observations yg and the observations of the regulators in πg. The time lag parameter τ describes how to shift the observations of the regulators in eq. (4), i.e., τ describes how to build the regressor matrices Xπg from the data D. Although Xπg depends on τ, we do not make this explicit in our notation.

As the targets g=1, …, G are modelled independently and the overall network structure ℳ is determined by the individual regulator sets, symbolically ℳ=(π1, …, πG), the joint marginal likelihood has a modular form:

(15)p(D|τ)=∏g=1G∫δg∑πg(∫wg∫σg2p(yg|Xπg, σg2, wg, πg, δg, τ)p(σg2)p(wg|σg2, δg, πg)dσg2dwg)p(πg)p(δg)dδg (15)

For a given regulator set πg and a fixed SNR-hyperparameter δg marginalizing the HBR likelihood over the regression parameters and the noise variances:

(16)p(yg|Xπg, πg, δg, τ)=∫σg2(∫wgp(yg|Xπg, σg2, wg, πg, δg, τ)p(wg|σg2, δg, πg)dwg)p(σg2)dσg2 (16)

yields a closed-form solution, see eq (15) in Aderhold et al. (2014).[6] It then follows from eqns. (15–16):

(17)p(D|τ)=∏g=1G∫δg(∑πgp(yg|Xπg, πg, δg, τ)p(πg))p(δg)dδg (17)

The prior, p(πg), on the regulator sets, πg, in the HBR model is assumed to be a uniform distribution subject to a constraint on the maximal cardinality ℱ of the number of regulators [see Grzegorczyk and Husmeier (2012) and Aderhold et al. (2014)]. We thus obtain:

(18)p(D|τ)=∏g=1G∫δg(1Tg∑πg:|πg|≤ℱp(yg|Xπg, πg, δg, τ))p(δg)dδg (18)

where Tg is the number of all valid regulator sets. Hence, if there are N potential regulators for a target g, then the number of valid parent sets Tg is given by:

(19)Tg=∑f=0ℱ(fN)=O(Nℱ) (19)

so that Tg grows polynomially with the power of ℱ.[7] If the inner sums can be computed by full-enumeration, the marginal likelihood can be estimated by repeatedly sampling δg (g=1, …, G) from their inverse Gamma prior distributions and computing the following Monte-Carlo estimator:

(20)p(D|τ^)=∏g=1G(1I∑i=1I(1Tg∑πg:|πg|≤ℱp(yg|Xπg, πg, δg(i), τ))) (20)

where δg(i) (g=1, … G; i=1, …, I) is the i-th sample drawn for target g from an inverse Gamma prior.[8] Alternatively, and in particular if the inner sums in eq. (18) are not computationally feasible, one can resort to standard numerical procedures based on MCMC, like Chib’s method (Chib and Jeliazkov, 2001).

To get an idea of the approximation error of our marginal likelihood estimator, we consider J independent Monte-Carlo estimators ψ^τ, 1,…,ψ^τ, J of size I, each computed with eq. (20). For each lag τ we can then consider the distributional form of these estimators to get an impression of (i.e., to bound) the approximation error. For the HBR model we have provided the technical details above, and we note that the marginal likelihoods of the HBR-nl and the HBR-light model can also be approximated along these lines. As we found that the marginal likelihoods of these three models show the same trends and peak at the same lag (a time shift of τ=2 h, referring to one single data point shift), we show only the results for the HBR model in Figure 8.

In our application the data set D consists of a set of individual time series. When the network interactions are modelled subject to a time lag corresponding to τ data points, then the first τ target observations have to be withdrawn from each time series.[9] For a fair comparison among different time lags τ we first choose a maximal lag τMAX, and we withdraw the first τMAX observations from all time series. Subsequently, the remaining target observations yg (g=1, …, G) can be used for all lags τ∈{0, …, τMAX}, i.e., this approach ensures that the target observations do not differ from τ to τ and that exactly the same target observations have to be modelled for all lags τ.[10]

Detailed simulation results

For our performance evaluation on the simulated data described in Section 4.1, we were running hundreds of simulations for a variety of different settings, related to the observation status of the molecular components (mRNA only versus mRNAs and proteins), the method for derivative estimation (described in Section 3.1), the regulatory network structure (shown in Figure 1), and the method applied for learning this structure from data (reviewed in Sections 3.2 and 3.3). The results from these studies are shown in Figures 12–14. These results are complex, and patterns are not easily discernible by visual inspection. This has motivated the application of the ANOVA scheme described in Section 3.5.

Figure 12 Detailed AUROC scores: coarse gradient.The boxplots show a fraction of the results obtained from the simulated data described in Section 4.1. The AUROC scores were obtained from the coarse response gradients with 2-h intervals. The corresponding results for the fine gradient and the Gaussian process interpolation are displayed in Figures 13, 14. Left panel: Incomplete data, with mRNA but no protein concentrations. Right panel: Complete data that include both protein and mRNA concentrations. Each panel contains six subpanels, representing the six different network topologies shown in Figure 1.
Figure 12

Detailed AUROC scores: coarse gradient.

The boxplots show a fraction of the results obtained from the simulated data described in Section 4.1. The AUROC scores were obtained from the coarse response gradients with 2-h intervals. The corresponding results for the fine gradient and the Gaussian process interpolation are displayed in Figures 13, 14. Left panel: Incomplete data, with mRNA but no protein concentrations. Right panel: Complete data that include both protein and mRNA concentrations. Each panel contains six subpanels, representing the six different network topologies shown in Figure 1.

Figure 13 Detailed AUROC scores: fine gradient.This figure corresponds to Figure 12, but shows the AUROC scores obtained with the fine gradient (24-min intervals) rather than the coarse gradient (2-h intervals). See Figure 12 for details.
Figure 13

Detailed AUROC scores: fine gradient.

This figure corresponds to Figure 12, but shows the AUROC scores obtained with the fine gradient (24-min intervals) rather than the coarse gradient (2-h intervals). See Figure 12 for details.

Figure 14 Detailed AUROC scores: fine gradient.This figure corresponds to Figure 12, but shows the AUROC scores obtained with the gradient from the Gaussian process interpolation rather than the finite difference method. See Figure 12 for details.
Figure 14

Detailed AUROC scores: fine gradient.

This figure corresponds to Figure 12, but shows the AUROC scores obtained with the gradient from the Gaussian process interpolation rather than the finite difference method. See Figure 12 for details.

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Published Online: 2015-2-14
Published in Print: 2015-4-1

©2015 by De Gruyter

Artikel in diesem Heft

  1. Frontmatter
  2. Research Articles
  3. Study of triplet periodicity differences inside and between genomes
  4. H-CLAP: hierarchical clustering within a linear array with an application in genetics
  5. Inferring bi-directional interactions between circadian clock genes and metabolism with model ensembles
  6. Bayesian inference for Markov jump processes with informative observations
  7. Likelihood free inference for Markov processes: a comparison
  8. Spatio-temporal model for multiple ChIP-seq experiments
  9. Software and Application Note
  10. GenePEN: analysis of network activity alterations in complex diseases via the pairwise elastic net
  11. Corrigendum
  12. Corrigendum to: Simple estimators of false discovery rates given as few as one or two p-values without strong parametric assumptions
https://doi.org/10.1515/sagmb-2014-0041
Schlagwörter für diesen Artikel
circadian clock; determination of time delays; model ensembles; plant metabolism; regulatory network inference

Artikel in diesem Heft

  1. Frontmatter
  2. Research Articles
  3. Study of triplet periodicity differences inside and between genomes
  4. H-CLAP: hierarchical clustering within a linear array with an application in genetics
  5. Inferring bi-directional interactions between circadian clock genes and metabolism with model ensembles
  6. Bayesian inference for Markov jump processes with informative observations
  7. Likelihood free inference for Markov processes: a comparison
  8. Spatio-temporal model for multiple ChIP-seq experiments
  9. Software and Application Note
  10. GenePEN: analysis of network activity alterations in complex diseases via the pairwise elastic net
  11. Corrigendum
  12. Corrigendum to: Simple estimators of false discovery rates given as few as one or two p-values without strong parametric assumptions
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