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In contrast to feature selection and gene set analysis, bi-level selection is a process of selecting not only important gene sets but also important genes within those gene sets. Depending on the order of selections, a bi-level selection method can be classified into three categories – forward selection, which first selects relevant gene sets followed by the selection of relevant individual genes; backward selection which takes the reversed order; and simultaneous selection, which performs the two tasks simultaneously usually with the aids of a penalized regression model. To test the existence of subtype-specific prognostic genes for non-small cell lung cancer (NSCLC), we had previously proposed the Cox-filter method that examines the association between patients’ survival time after diagnosis with one specific gene, the disease subtypes, and their interaction terms. In this study, we further extend it to carry out forward and backward bi-level selection. Using simulations and a NSCLC application, we demonstrate that the forward selection outperforms the backward selection and other relevant algorithms in our setting. Both proposed methods are readily understandable and interpretable. Therefore, they represent useful tools for the researchers who are interested in exploring the prognostic value of gene expression data for specific subtypes or stages of a disease.
Feature selection, where the primary objective is selection of individual relevant variables, such as genes associated with a phenotype of interest, is an important topic in the field of bioinformatics. Because gene expression profiles exhibit grouped structure with genes inside each group being highly correlated to each other, gene set analysis has become popular and widespread over the last decade1,2. The goal of a gene set analysis method is to examine the association of a gene set with the phenotype of interest, meaning the selection of relevant gene sets. Both gene and gene set selections lead to more parsimonious final models with better predictive performance and more meaningful biological interpretation.
In contrast to feature selection and gene set analysis, bi-level selection is a process of selecting relevant features at two levels, as its name implies3. Bi-level selection is motivated by the fact that some genes within a gene set may be unrelated to the phenotype of interest, although the gene set as a whole is involved in the biological process. Since a bi-level selection analysis is interested in selecting not only important groups/gene sets but also those important members/genes inside those groups/gene sets, a feature may be thus referred to as either a gene set or a gene here.
Based on the order of selection on genes and gene sets, bi-level selection methods can be classified into three categories: (1) the forward selection method which first selects relevant gene sets followed by the selection of relevant individual genes, e.g.,4; (2) the backward selection method such as5,6,7 which takes the reversed order; and (3) the simultaneous selection method which selects important gene sets and genes at the same time, usually by the means of adding a penalty term which penalizes all genes inside the same group similarly, e.g.,3,8. The simultaneous selection method using a penalized regression model is more statistically sophisticated, compared to the first two methods. Here we focus on the forward and backward selection methods due to their ease in implementation and thus wide utilization.
Even though both the forward and backward selection methods can implement two level selections, their relative emphases on these two levels may differ. Nevertheless, their common drawback is that the small number of selected features in the first step may lead to challenges in performing the second step. Intuitively, the forward method is superior to the backward method because the extraction of core genes before the identification of important gene sets may screen out the gene sets whose members have subtle individual effects but their coordinated effects being significant when taken together.
Non-small cell lung cancer (NSCLC) is a leading cause of cancer deaths in many countries9. Adenocarcinoma (AC) and squamous cell carcinoma (SCC) are two major histological subtypes of NSCLC. There is increasing evidence to support that AC and SCC differ in the composition of genes and molecular characteristics10. Therefore, they should be regarded as two distinct diseases and treated with different treatment strategies.
Currently, treatment choices for these two subtypes are very homogeneous, mainly depending on the stage at which the cancer is diagnosed. It is becoming critical to evaluate the risk profiles of patients using a reliable gene/gene set signature. Given the fundamental differences between AC and SCC, the genes/gene sets associated with recurrence and survival rates for each histology subtype are expected to be different11,12,13. To test the existence of subtype-specific prognostic genes, two statistical methods were proposed previously by us11,12. For example, the so-called Cox-filter model11 fits a Cox model using gene, subtype, and their interaction term as covariates for each gene. After filtering and excluding insignificant genes, the Cox-filter model identifies subtype-specific prognostic genes (discussed in detail in the Methods section). However, the Cox-filter method does not take the grouping structures among genes into account, and thus selects genes based on the strength of their individual effects and introduces false positives.
In this study, we extend the Cox-filter model to two bi-level selection methods — one forward method and one backward method — by constructing patients’ risk profiles at the gene set level using the sign average method6,14. The two Cox-filter extensions are then applied to NSCLC data to investigate if they can alleviate the limitations of the original Cox-filter model.
To examine the characteristics of our proposed procedures, and to explore if both extensions can alleviate the disadvantages of the Cox-filter method, we used actual expression values of the microarray data to conduct simulations. We used the first 100 gene sets in the c5 category of the Molecular Signatures Database and then randomly selected 4 genes – ARRB1, COPA, ECE2, and SMAD4 – to be prognostic markers. Among them, COPA, ECE2 and SMAD4 belong to the same gene set while ARRB1 is inside a different gene set.
Similar to the simulation setting in our previous study, we considered two extreme cases: (1) the mutually exclusive prognostic genes case in which AC and SCC have completely distinct sets of survival-relevant genes, and (2) no subtype-specific prognostic genes case in which AC and SCC share identical prognostic markers. In other words, for the first case, βACs for genes 1 and 3 while βsccs for genes 2 and 4 are set to non-zero values. The hazard functions are specified as:
the survival time for each patient was simulated via a Cox-exponential distribution15 according to the above hazard functions, and the censoring rate was set at 30%.
For the other case, the hazard function is identical for both AC and SCC subtypes:
this means that all βACs and βsccs for genes 1–4 are expected to have non-zero coefficients. For both scenarios, we simulated 50 datasets and applied our proposed procedures to these replicates. The percentages of the causal genes being selected over the 50 replicates are summarized in Table 1 for simulation 1 and in Table 2 for simulation 2, respectively. In both tables, the C-statistics and the Rand Index at the gene level and their standard errors are also provided, on which the comparison between the bi-level selection methods and other relevant algorithms, namely, the Cox-filter model11, the Cox-TGDR method12, and a separate LASSO16 for each subtype, was made.
In both simulations, the forward Cox-filter was superior to the backward Cox-filter in terms of the frequencies of identifying the true causal genes and the sizes of the final models. Compared with the original Cox-filter method, which selects genes simply based on individual effects, both the forward Cox-filter method and the backward Cox-filter method are more likely to select those genes with small individual effects but significant coordinated effects together with other genes in the same gene set, i.e., COPA, ECE2, and SMAD2. Nevertheless, as a filter method to select genes, the model parsimony (i.e., the number of genes in the resulting prognostic gene signature) of these two proposed methods remains unsatisfactory.
As a reference, the percentages of these four causal genes being selected under a random guess model by both forward Cox-filter and backward Cox-filter are presented in Table 3. Under the null model, almost all percentages of these genes being selected are zeros, suggesting the high frequencies obtained by both bi-level selection methods are impossible to occur by chance alone.
Overall, the forward Cox-filter method has the best performance when the C-statistics and the Rand Index are considered together. The backward Cox-filter method shows no such an overall superiority. For instance, in the first simulation the backward Cox-filter method has the lowest Rand index (15.87%) among these 5 algorithms but a very large standard error (5.04%) for the SCC subtype.
In this application, we first trained on the NSCLC microarray data using both proposed methods and validated the predictive performance of the resulting gene signatures using the RNA-seq data as a test set. Then we reversed the order of the training set and the test set. Lastly, we applied three relevant methods—the Cox-filter method and the Cox-TGDR method proposed by us11,12 to test the existence of subtype-specific prognostic genes, and LASSO for survival analysis17 – to the NSCLC microarray data and compared both proposed bi-level selection methods and the three relevant methods. The performance statistics, i.e., the C-statistics calculated by applying the prognostic signatures on the test set and the Rand Index at the gene and gene set levels using 10-fold cross-validations, and their standard errors obtained using bootstrapped samples are presented in Table 4.
The forward Cox-filter method outperformed the three other relevant feature selection methods and the backward Cox-filter method. Of note, a non-statistically different C-index value of 52.8% (standard error: 2.75%) from a random guess model was obtained by the backward selection method for the SCC subtype. It is consistent with our expectation that the performance of the backward selection method is inferior to that of the forward selection method and the simulation results. However, we don’t exclude the likelihood that the backward selection method may be optimal for some specific data types or structures.
To verify if the resulting prognostic signatures are confounded by other known variables, i.e., age, sex, and smoking status, a multiple Cox regression model was fitted using the risk scores estimated from the resulting prognostic signature and these clinical variables as covariates. The results are presented in Table 5. Based on the most significant p-values for the prognostic signatures, it is concluded that the resulting prognostic signatures possess adequate prognostic capacity in predicting the survival of NSCLC patients.
Moreover, as shown by the Venn-diagrams in Fig. 1, the selected gene sets and genes using the forward method and the backward method share no or limited overlap. This finding indicates that the focuses of the two methods might be distinct. While for the NSCLC application both methods tend to improve the pathway-level and gene-level stabilities, it appears that the increment in pathway-level stability for the forward Cox-filter method is dramatically larger than the gene-level stability. In contrast, the backward Cox-filter method does not possess this feature. Such a pattern has been overlooked by previous work in which researchers only illustrate when a method accounts for pathway knowledge, its stabilities at both gene and gene set levels may be improved.
In contrast, the overlaps between respective gene sets and genes for AC and SCC using either the forward method or the backward method are substantially larger, implying that there might exist more pan genes or gene sets for both subtypes than subtype-specific ones. Although we emphasize the importance of those subtype specific gene/gene sets, the critical role played by those pan gene/gene sets cannot be denied as well.
Using simulated data and a real-world application, we demonstrate that the forward Cox-filter method outperforms relevant algorithms under consideration as well as its backward counterpart in terms of gene-level and pathway-level stabilities and performance statistics. Given that there are numerous pathway analysis methods and feature selection algorithms, the forward Cox-filter method cannot be the “supermodel”—the optimal model for every expression data. Furthermore, as a bi-level selection method, the performance of the forward Cox-filter method relies on the quality of pathway knowledge. More specifically, these two proposed bi-level selection methods make an implicit assumption that all genes within a specific gene set shall co-function together to influence the phenotype of interest, which may not be true for all diseases.
We emphasize that we do not make any recommendation on the clinical utilization of the resulting prognostic gene signatures or gene set signatures. The primary objective of this study is to introduce the two bi-level selection methods. These two methods are easily implementable and readily interpretable, even for a biologist or clinician since it builds upon the Cox-models18 and the signed averages of expression values over a gene set or selected subset14. Therefore, they are easily accessible for researchers who are interested in exploring the prognostic value of gene expression data for specific subtypes or stages of a disease. As the first effort to address the issue of identifying both subtype-specific prognostic gene sets and genes for the early-stage NSCLC patients while accounting for the pathway information, the proposed methods may spark interest in this research area and propel the development of more advanced statistical methods.
The RNA-Seq data for those patients at early histology stages (stages I and II) were downloaded from The Cancer Genome Atlas (https://tcga-data.nci.nih.gov/tcga/). By restricting the patients to those at early stages and being adjuvant treatment naïve with survival information, 70 AC and 55 SCC subjects remained.
The microarray data used were the experiments of GSE50081 in the GEO repository. The chips in this dataset were hybridized on the Affymetrix HGU133Plus 2.0 platform. After deleting those samples with ambiguous labels, 127 AC and 42 SCC patients were included in this study.
Raw data (CEL files) of the microarray data set were downloaded from the GEO repository. Expression values were obtained using the fRMA algorithm21, and were normalized using quantile normalization. Then moderated t-tests using limma22 were carried out to identify the differentially expressed genes (DEGs) between SCC and AC in the microarray data set, and those non-DEGs with the false discovery rate (FDR)>0.05 were filtered out. To deal with multiple probe sets matched to one specific gene, the one with the largest fold change was retained.
For the RNA-seq data, Counts-per-million (CPM) values were calculated and log2 transformed by the Voom function23 in R limma package22. The downstream analysis was conducted upon the 2569 genes inside microarray data, RNA-seq data, and the annotated gene sets.
The Cox-filter method proposed by us11 is used to identify genes informative of survival rate for AC/SCC histology subtypes. In this method, a Cox model is fitted on each gene, and the hazard function of patient i for gene g (g=1,…,p) at time point t is given by,
here, Xij=(Xij1,…,Xijp)T represent expression values for the p genes under consideration and λ0g(t) is an unknown baseline hazard function at time point t. I(j=SCC) is an indicator, taking the value of 1 if the histology subtype j of patient i is SCC. Otherwise, its value is 0 if the histology subtype of this patient is AC. Both β2g and β3g are the parameters of interest, with β2g representing the change in log hazard rate associated with 1-unit increase in the actual expression value of gene g among AC, and β3g representing the additional change in log hazard rate associated with the SCC subtype. The values of βACg, i.e., β2g, and βSCCg, i.e., β2g+β3g, determine if subtype-specific prognostic genes do exist. For example, βACg≠0 but βSCCg=0 corresponds to an AC-specific gene and βSCCg≠0 but βACg=0 corresponds to an SCC-specific gene. In practice, we may also be interested in the scenario of both βACg and βSCCg having different non-zero values.
After fitting the Cox-filter model for each gene and obtaining estimated βACg and βSCCg for each gene, we take the membership of genes into consideration and then summarize a patient’s risk profile as the sign average of his/her expression values over all genes/selected genes inside each specific gene set.
Specifically for each subtype, all genes/selected genes inside this gene set are classified into two groups according to the signs of their estimated effects in the above equation — the hazardous group H and the preventive group P. In the first group, genes with increased expression that are associated with higher hazard are included. In contrast, genes for which an increment in expression reduces hazard are classified into the second group. Since one gene may be hazardous in one subtype while preventive in the other, we introduce two sets of notations for the AC and SCC patients, i.e., HAC and PAC for AC patients, and HSCC and PSCC for SCC patients.
Then the sign average Zijk for patient i of subtype j (either AC or SCC) in gene set k is calculated as
here is the number of genes inside gene set . and Xil represents the gene expression value of gene l (l includes all genes belong to one specific gene set) for patient i. By taking the difference between the sum of expression values of all genes inside the hazardous group H and that of all genes inside the preventive group P and then dividing by the total number of genes/selected genes, Zijk is essentially the signed average of expression values over genes under consideration inside gene set k.
In order to implement bi-level selection with the Cox model in Eq. 3, we apply it twice in a either forward or backward way. Correspondingly, we refer them as the forward Cox-filter method and the backward Cox-filter method.
In the forward Cox-filter extension, the risk profiles for patient i are first calculated over all genes inside the specific gene set k, denoted as Zijk. Here, Zij=(Zij1,…,ZijK)T is a K-length vector representing the gene set level risk profiles for patient i over K gene sets under consideration.
Replacing genes by gene sets, then the Cox-filter model is fitted again. For one specific gene set k (k=1,2,…,K, here K is the number of gene sets under consideration), the Cox-filter model may be expressed as,
where Zijk is the sign average obtained in Eq.4, representing the expression level of gene set k for patient i. After filtering out those insignificant gene sets for whose adjusted p-values (using the procedure) are larger than a pre-determined cut-off, the significant levels of genes inside the identified gene sets are determined on the basis of the adjusted p-values of the Cox-filter models in Eq. 3.
Taking the reversed selection orders, the backward Cox-filter extension first selects candidate genes based on the adjusted p-values of the Cox-filter models in Eq. 3 and then calculates the risk profiles over those selected genes for each gene set. Lastly, the significant levels of gene sets are determined using the adjusted p-values of the Cox models in Eq. 5. Correspondingly, the final selected genes are those involved in the first step and also contained inside those significant gene sets.
In both procedures, the adjusted p-values in Eq. 3 and Eq. 5 may be treated as tuning parameters. Over a grid of values, i.e., 0.01, 0.05, 0.1, 0.15 and 0.2, their optimal values are decided using 10-fold cross-validations. Figure 2 provides graphical elucidation of both the forward Cox-filter method and the backward Cox-filter method.
We use the censoring-adjusted C-statistic24 over the follow-up period (0,τ) to evaluate the performance of a resulting prognostic gene signature, where τ is a pre-specified time point. The C-statistic is defined as,
where g(Xi) is the risk score for subject i with predictor vector Xi and Ti is the survival time of patient i.
The C-statistic can be interpreted as the probability of concordance between the predicted and observed survival times over all pairs of observations over the follow-up period (0,τ). Its asymptotic distribution is presented in the Appendix of the original paper24. Empirically, even though a value of 0.5 for the C-statistics corresponds to the random guess model, a prognostic signature with the C-statistic of relatively moderate values, i.e., between 0.6–0.7 is regarded to have satisfactory predictive performance25. The calculation of the C-statistic is implemented using the R survAUC package.
In addition, the Rand index is calculated to evaluate the stability or robustness of the resulting signatures. With k runs (e.g., the runs in an k-fold cross-validation or the applications of the method to k different data sets) of an algorithm, k lists of genes are obtained, i.e., gs1, gs2, … gsk. Then a Rand index is defined as
where ∩ represents the intersection between two gene lists, represents the union between the gene lists gsi and gsj, and | | stands for the size of the corresponding set. The Rand index represents the agreement among the resulting signatures trained from different data sets.
With the gene lists being replaced by the pathway lists, the stability of resulting gene sets is evaluated using the Rand index.
All statistical analysis was carried out in the R language version 3.2 (www.r-project.org).
How to cite this article: Tian, S. et al. Identification of prognostic genes and gene sets for early-stage non-small cell lung cancer using bi-level selection methods. Sci. Rep. 7, 46164; doi: 10.1038/srep46164 (2017).
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This study was supported by a fund (No. 31401123) from the Natural Science Foundation of China. We thank Dr. Catherine Anthony for English editing.
The authors declare no competing financial interests.
Author Contributions Conceived and designed the study: S.T., J.S. Analyzed the data: S.T., C.W. Interpreted data analysis and results: S.T., H.H.C., J.S., C.W. Wrote the paper: S.T., H.H.C. All authors reviewed and approved the final manuscript.