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Reproducible detection of disease-associated markers from gene expression data
- Katsuhiro Omae^{1}Email author,
- Osamu Komori^{2} and
- Shinto Eguchi^{1, 3}
https://doi.org/10.1186/s12920-016-0214-5
© The Author(s) 2016
Received: 3 August 2015
Accepted: 3 August 2016
Published: 18 August 2016
Abstract
Background
Detection of disease-associated markers plays a crucial role in gene screening for biological studies. Two-sample test statistics, such as the t-statistic, are widely used to rank genes based on gene expression data. However, the resultant gene ranking is often not reproducible among different data sets. Such irreproducibility may be caused by disease heterogeneity.
Results
When we divided data into two subsets, we found that the signs of the two t-statistics were often reversed. Focusing on such instability, we proposed a sign-sum statistic that counts the signs of the t-statistics for all possible subsets. The proposed method excludes genes affected by heterogeneity, thereby improving the reproducibility of gene ranking. We compared the sign-sum statistic with the t-statistic by a theoretical evaluation of the upper confidence limit. Through simulations and applications to real data sets, we show that the sign-sum statistic exhibits superior performance.
Conclusion
We derive the sign-sum statistic for getting a robust gene ranking. The sign-sum statistic gives more reproducible ranking than the t-statistic. Using simulated data sets we show that the sign-sum statistic excludes hetero-type genes well. Also for the real data sets, the sign-sum statistic performs well in a viewpoint of ranking reproducibility.
Keywords
Background
Detection of disease-associated markers plays a crucial role in gene screening for biological studies. In this field, statisticians seek to identify informative genes as candidates for further investigation. To this end, it is desirable to correctly rank genes according to their degree of differential expression. In such efforts, two-sample test statistics, such as the t-statistic and Wilcoxon sum-rank statistic, are widely used to rank genes based on gene expression data.
Such heterogeneity was also discussed in a context of cancer outlier [4, 5] which developed the methods using the two-sample test statistic to detect genes that are over or down expressed in the subset of the disease group compared with the normal group. In this paper, we focus not on the subset but rather on the whole set. That is, our main aim is to detect the genes that are differentially expressed entirely in the disease group. Thus we develop the ranking method which has robustness for the heterogeneity.
Novel methods such as lasso are able to determine both highly ranked genes and classifiers simultaneously. However, [6] supported the importance of choosing a filtering method that yields a gene ranking corresponding to feature selection, rather than to the classification in machine-learning theory. In other words, pre-selection and evaluation of the resultant gene set must be separated from the classifier’s performance. Moreover, each top-ranked gene in itself must be informative or effective in some sense, e.g., robustness with respect to heterogeneity focused in this paper. Therefore we consider the ranking score independently derived for each gene, unlike [7], which took correlations between genes into consideration.
Because t-statistics and correlations strongly fluctuate due to sample variation, combining a sampling method with a two-sample test statistic should improve reproducibility. The effects of sampling method have been demonstrated by multiple studies, both theoretical [8] and applied [9]. Meanwhile, [10] presented counterarguments: several approaches to feature selection with ensemble learning by the sampling method are ineffective in terms of predictive ability, stability, and interpretability. Those authors concluded that the simple Student’s t-test exhibits superior performance in these regards. Dabney [11] also argued that the simple t-statistic is more accurate than the modified t-statistic and shrunken centroids. However, those authors’ conclusions are based on empirical studies in the context of particular data. We argue that the sampling method is effective from the standpoint of robustness with respect to heterogeneous factors; this is because heterogeneity represents a mixture of two or more classes, and we can easily imagine that sampling is the best way to capture heterogeneity to integrate information from many small subsample sets. To stabilize the performance of the simple t-statistic, we derived a sign-sum statistic that improves ranking reproducibility. This novel statistic repeatedly counts the sign of mean difference between subsets of the normal and disease groups. The sign-sum statistic is an extension of the Wilcoxon sum-rank statistic, which itself has superior robustness but an inferior power relative to the t-statistic [12]. We show the probabilistic result of the sign-sum statistic, and demonstrate its superior performance through simulations and applications to real data in this paper.
Methods
Derivation
where \(\bar {X}_{jy}\) is the sample mean of group y for gene j. There are two options for s _{ j }, a pooled Student’s type or a non-pooled Welch’s type. In this paper, we use Welch’s t-statistic; therefore, s _{ j } is written as \(s_{j} = \sqrt {s_{1j}^{2} / \hat {\pi }_{1}+s_{0j}^{2}/ \hat {\pi }_{0}}\), where s _{ yj } is the sample standard deviation of gene j and \(\hat {\pi }_{y} = n_{y} / n\) for group y. Without loss of generality, we can assume that \(\bar {X}_{j1}-\bar {X}_{j0}\geq 0\).
However, as discussed in Background, the t-statistics can fluctuate between two divided data sets, and even the signs of t-statistics can be mismatched. Therefore, we focus our attention to the signs of the t-statistics. If the t-statistics are evaluated by the full sample, the signs are positive over all genes by the assumption above. However, if the t-statistics are evaluated by subsets of the full sample, the signs may change, as shown in Fig. 1. Therefore, we derived a sign-sum statistic to count the signs of the t-statistics for all possible subsets.
where \(k_{1}=\binom {n_{1}}{a}, k_{0}=\binom {n_{0}}{b}, \mathrm {H}(x)\) is a Heaviside-step function that takes the value 0 if x<0 or 1 otherwise, and \(\bar {X}_{jyt}\) is the sample mean of gene j in the t-th subset of group y. A larger value of the sign-sum statistic means that the signs of the t-statistics evaluated by subsamples are more stably positive. We can show that the sign-sum statistic is an extension of Wilcoxon’s sum-rank statistic; in fact, if a and b are equal to 1, then they are equivalent.
These statistics are described by the character U because they are members of t-statistics, as shown in Additional file 1. We compare the sign-sum statistic (2) with the t-statistic evaluated by subsamples (3) from the perspective of t-statistics in the next subsection.
By the assumption \(\bar {X}_{j1}-\bar {X}_{j0}\geq 0\), the gene which has a larger score of the statistic is regarded as more informative for the detection of differentially expressed genes; thus, the gene ranking is obtained by sorting the values of the statistics in descending order over all genes.
Robustness for heterogeneity
Heterogeneous disease factors can cause a mixture of two or more classes in some gene expression levels in the disease group. We call genes affected by such factors as “hetero genes”, and unaffected genes as “homo genes”. The sign-sum statistic can effectively detect such heterogeneity. To demonstrate this, here we provide a theorem about asymptotic confidence intervals.
Let U be a general two-sample U-statistic (we drop the gene index for simplicity). Because the U-statistic has the property of asymptotic normality, the asymptotic confidence interval is described as \(\mathrm {E}[U] \pm (\sigma _{U} / \sqrt {n}) Z_{\alpha /2}\), where \({\sigma ^{2}_{U}}\) is an asymptotic variance of U and Z _{ α/2} is the 100α/2 upper percentile of a standard normal distribution. Because U ^{ T } and U ^{ S } are members of U-statistics, these statistics are evaluated by the interval estimators as shown in Theorem 1.
Theorem 1
- 1.The asymptotic confidence interval of the t-statistic evaluated by subsamples with level α is$$\begin{array}{@{}rcl@{}} \frac{\sqrt{a+b} \ (\mu_{1}-\mu_{0})}{\sqrt{\cfrac{{\sigma_{1}^{2}}}{\pi_{1}}+\cfrac{{\sigma_{0}^{2}}}{\pi_{0}}}} \pm Z_{\alpha/2} \frac{(a+b)^{1/2}}{\sqrt{n}} \end{array} $$(4)
if \(\hat {\pi }_{1} \rightarrow \pi _{1}\) and \(\hat {\pi }_{0} \rightarrow \pi _{0}\), where π _{1}+π _{0}=1 and Z _{ α/2} is 100α/2 upper percentile of a standard normal distribution.
- 2.The asymptotic confidence interval of the sign-sum statistic (2) with level α is$$\begin{array}{@{}rcl@{}} \mathrm{E}[U^{S}] \pm Z_{\alpha/2} \frac{\tilde{\sigma}}{\sqrt{n}}, \end{array} $$(5)if \(\hat {\pi }_{1} \rightarrow \pi _{1}\) and \(\hat {\pi }_{0} \rightarrow \pi _{0}\), where Z _{ α/2} is 100α/2 upper percentile of a standard normal distribution, and$$\begin{array}{@{}rcl@{}} \mathrm{E}[U^{S}]&=&\mathrm{E}[G_{1}(V_{1})], \end{array} $$(6)$$\begin{array}{@{}rcl@{}} \tilde{\sigma}^{2} &=&\frac{a^{2}}{\pi_{1}}\text{Var} [G_{1} (V_{1})]+\frac{b^{2}}{\pi_{0}}\text{Var} [G_{0} (V_{0})], \end{array} $$(7)where G _{ y }(v)=Pr(W _{ y }≤v) for y=0,1, and$$\begin{array}{@{}rcl@{}} V_{1} &=& \frac{1}{a}X_{11}, W_{1} =-\frac{1}{a}\sum\limits_{i=2}^{a} X_{1i} + \frac{1}{b} \sum\limits_{j=1}^{b} X_{0j}, \\ V_{0} &=& -\frac{1}{b}X_{01}, W_{0} =-\frac{1}{a}\sum\limits_{i=1}^{a} X_{1i} + \frac{1}{b} \sum\limits_{j=2}^{b} X_{0j}. \end{array} $$
Here X _{1}s and X _{0}s are independently distributed with F _{1} and F _{0}, which denote the distribution functions of gene expression levels of the disease and normal groups, respectively.
A proof of the Theorem 1 is given in Additional file 1. We note that V _{1}−W _{1} and V _{0}−W _{0} represent the mean differences in the disease and normal groups, respectively. A property of U-statistics allows us to evaluate the asymptotic variance of the sign-sum statistic by the conditional distribution of W _{ y } given V _{ y } for each group y. The difference between these two statistics is mainly due to the fact that the sign-sum statistic is the sum of the non-linear functions of the t-statistic evaluated by subsamples. As a result, information about F _{1} and F _{0} is strongly reflected in the sign-sum statistic as a result of changing a and b. We can discriminate hetero genes from homo genes by this property, as shown in the next subsection.
The effects of different setting of parameters
The difficulty and importance of considering such hetero genes is also discussed in [13] in the context of the false positive rate. The sign-sum statistic repeatedly counts the sign of the difference between the means of the disease and normal groups. Hence, the sign mismatches due to heterogeneity in the disease group would be effectively detected by a small a value, chosen such that the sample mean of the disease group fluctuates. This consideration is supported through numerical evaluations of specific situations, as described below.
where \(\pi _{1}, \pi _{0}, \mu _{1}^{*},\) and \(\sigma _{1}^{*}\) are the sample ratio of the disease group, the sample ratio of the normal groups, the expected mean, and the expected standard deviation of the hetero gene. The ranking by t-statistics fluctuates because the interval estimators of the hetero and homo genes are almost overlapping.
Simulation
We carried out simple simulation studies to evaluate the performance of the sign-sum statistic. With the number of genes set at 1000, we generated expression levels for 100 homo genes and 100 hetero genes; the remaining 800 were non-informative genes whose expression level distributions were equal in the disease and normal groups. Gene expression levels in the normal group were assumed to be drawn from a standard normal distribution N(0,1) without loss of generality. Homo gene expression levels in the disease group were drawn from a normal distribution \(N(1,{\sigma _{1}^{2}})\), and hetero gene expression levels in the disease group were drawn from a normal mixture distribution τ _{1} N(0,1)+τ _{2} N(m _{2},1), where τ _{1},τ _{2} are positive values with τ _{1}+τ _{2}=1. The mixture model suggests that a proportion τ _{1} of gene expression levels in the disease group cannot be discriminated from those in the normal group, as in real data.
We considered three situations in which the t-statistic confuses the homo and hetero genes by the constraint as (9) with different parameters: (I) \({\sigma _{1}^{2}}=1, \tau _{1}=0.5\), (II) \( {\sigma _{1}^{2}}=4, \tau _{1}=0.75\) and (III) \({\sigma _{1}^{2}}=1, \tau _{1}=0.25 \), with sample size n=200,1000 with equal n _{0} and n _{1}. We compared the gene rankings among three statistics: simple t-statistic with subsamples, simple t-statistic without subsamples, and sign-sum statistic with 100 repetitions. The sampling sizes were fixed as a=1 and b=1 for the t-statistic, and as a=1 and b=1,5 and 10 for the sign-sum statistic. Robustness with respect to heterogeneity was calculated based on the number of homo genes in the top 100 ranking. Although \({U_{j}^{T}}\) and \({U_{j}^{S}}\) are defined by all possible subsets, in this case we only need to evaluate sufficient combinations to achieve convergence of the top 100 rankings as written in Additional file 2.
Application
We compared the t-statistic with the sign-sum statistic using five real data sets [2, 14–17]. The data set in [2] (breast cancer data) contains 97 gene expression subjects for primary breast tumors in which 46 subjects are in relapsed group and 51 subjects are in relapse-free group for 5 years. We applied the same filtering used in [2], yielding a final full data set consisting of 97 samples and 5420 genes. The data set in [14] (cohort data) combine 454 gene expression samples from different diseases. We picked 32 samples from lung cancer tumors, 45 samples from pancreatic ductal adenocarcinoma tumors, and 70 samples from unaffected individuals, yielding a final full data set consisting of 147 samples and 863 genes. The data set in [15] (prostate cancer data) contains 6144 gene expressions for 455 prostate cancer tumors in which 103 subjects are determined as fusion status-positive and 352 subjects are determined as fusion status-negative. The data set in [16] (breast cancer data2) contains 17489 gene expressions for 286 breast cancer tumors in which 107 subjects are in relapsed group and 179 subjects are in relapse-free group within 5 years. The data set in [17] (leukemia data) contains 7129 gene expressions for 72 leukemia samples in which 47 subjects are in acute lymphoid leukemia group and 25 subjects are in acute myelogenous leukemia group.
where \(N_{p,k}=\sum _{i=0}^{k} i \binom {k}{i} \binom {p-k}{k-i}/\binom {p}{k}{=k^{2} / p}, S_{1t}\) and S _{2t } are the top k-ranked genes sets for two divided data on the t-th trial; in this case, k=100 and T=100. N _{ p, k } refers to the expected overlap in gene number for a random selection. A larger ORRS value means that the selection is more reproducible than the random selection.
Results and discussion
The performance of the sign-sum statistic
The number of homo genes in the top 100 ranking: these are obtained using the t and sign-sum statistics. Means(sd) from 100 repetitions for each situations and sample size is written
n=200 | |||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
Situation I | 50.0 (4.12) | 49.7 (3.67) | 61.7 (3.20) | 80.8 (2.56) | 83.1 (2.36) |
Situation II | 49.9 (3.50) | 49.3 (3.49) | 48.9 (3.19) | 56.5 (3.16) | 57.8 (3.23) |
Situation III | 49.7 (3.09) | 49.4 (3.31) | 72.1 (2.77) | 73.3 (3.12) | 72.4 (3.14) |
n=1000 | |||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
Situation I | 49.9 (4.19) | 49.9 (4.08) | 75.2 (2.98) | 97.3 (1.20) | 98.2 (1.00) |
Situation II | 50.1 (3.58) | 49.9 (3.86) | 47.6 (3.48) | 64.0 (3.07) | 67.2 (3.13) |
Situation III | 50.2 (3.65) | 50.3 (3.82) | 90.7 (2.19) | 92.6 (2.04) | 92.4 (1.98) |
Reproducibility and ORRS: these values indicate mean(sd) and were evaluated by 100 random separations of the full data
Reproducibility | t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} |
Breast cancer data | 3.78 (1.92) | 3.68 (1.99) | 4.33 (2.13) | 6.70 (2.77) | 7.39 (3.37) |
Cohort data | 23.7 (4.94) | 23.5 (4.86) | 27.5 (5.70) | 43.4 (5.39) | 42.6 (5.28) |
Prostate cancer data | 33.4 (4.29) | 32.6 (4.69) | 39.6 (4.76) | 31.4 (4.64) | 29.4 (4.23) |
Breast cancer data2 | 1.39 (1.35) | 1.42 (1.31) | 1.00 (1.10) | 3.33 (1.80) | 3.82 (1.91) |
Leukemia data | 32.3 (4.47) | 31.8 (4.41) | 34.2 (4.59) | 37.4 (4.30) | 37.1 (4.22) |
ORRS | t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} |
Breast cancer data | 2.20 (1.11) | 2.14 (1.16) | 2.52 (1.24) | 3.90 (1.62) | 4.30 (1.96) |
Cohort data | 2.02 (0.42) | 2.01 (0.42) | 2.35 (0.49) | 3.71 (0.46) | 3.63 (0.45) |
Prostate cancer data | 20.0 (2.57) | 19.5 (2.81) | 23.7 (2.85) | 18.8 (2.78) | 17.6 (2.54) |
Breast cancer data2 | 2.73 (2.64) | 2.78 (2.57) | 1.96 (2.16) | 6.53 (3.53) | 7.49 (3.75) |
Leukemia data | 19.3 (2.68) | 19.1 (2.64) | 20.5 (2.75) | 22.4 (2.57) | 22.2 (2.53) |
Discussion
Gene ranking procedures are not reproducible among different studies [18]. To obtain a robust ranking, ensemble or resampling methods are effective [8, 9]. Counterintuitively, however, resampling methods do not improve reproducibility [10]. In this paper, we evaluated a resampling method for robustness with respect to heterogeneity in a microarray study. We focused on the sign mismatch of t-scores in the context of a classification problem. We often found that the genes with large t-scores in the training data had small or sign-reversed t-scores in the test data. The sign-sum statistic was developed based on these two motivations. Using numerical simulation, we proved that the sign-sum statistic improves the robustness with respect to heterogeneity relative to the t-statistic. Furthermore, the sign-sum statistic allowed us to obtain a reproducible ranking in applications to real data. These conclusions were validated by an evaluation of the upper confidence limit (Theorem 1).
In the context of gene screening, FDR (False Discovery Rate) has been studied by novel methods such as SAM [19] and ranking procedure by q-values [20] for decisions about the cut-off value for gene ranking. It is less meaningful to focus on the cut-off value until we have a correct and stable gene ranking. Therefore, in this study, we focused on obtaining a reproducible gene ranking. Obtaining the cut-off value of the sign-sum statistic is a goal for future work.
Test AUC for four real data sets: each predictor is constructed by the DLDA rule. These values indicate mean(sd) and were evaluated by 100 random separations of the full data
breast cancer data | AUC of the test data by DLDA | ||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
10 genes | 0.698 (0.058) | 0.698 (0.058) | 0.698 (0.060) | 0.684 (0.065) | 0.679 (0.069) |
50 genes | 0.705 (0.046) | 0.705 (0.047) | 0.707 (0.050) | 0.712 (0.051) | 0.712 (0.050) |
100 genes | 0.711 (0.045) | 0.710 (0.045) | 0.712 (0.047) | 0.718 (0.045) | 0.718 (0.045) |
Cohort data | AUC of the test data by DLDA | ||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
10 genes | 0.744 (0.061) | 0.743 (0.061) | 0.751 (0.064) | 0.755 (0.064) | 0.771 (0.063) |
50 genes | 0.779 (0.057) | 0.778 (0.057) | 0.773 (0.053) | 0.784 (0.053) | 0.789 (0.054) |
100 genes | 0.782 (0.056) | 0.781 (0.057) | 0.778 (0.054) | 0.781 (0.052) | 0.784 (0.051) |
Prostate cancer data | AUC of the test data by DLDA | ||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
10 genes | 0.835 (0.025) | 0.835 (0.026) | 0.832 (0.025) | 0.823 (0.027) | 0.808 (0.032) |
50 genes | 0.846 (0.023) | 0.845 (0.024) | 0.847 (0.022) | 0.836 (0.029) | 0.829 (0.033) |
100 genes | 0.844 (0.024) | 0.842 (0.025) | 0.848 (0.021) | 0.829 (0.030) | 0.822 (0.033) |
Breast cancer data2 | AUC of the test data by DLDA | ||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
10 genes | 0.612 (0.044) | 0.614 (0.041) | 0.611 (0.043) | 0.595 (0.042) | 0.581 (0.044) |
50 genes | 0.634 (0.040) | 0.633 (0.040) | 0.630 (0.042) | 0.623 (0.039) | 0.619 (0.040) |
100 genes | 0.637 (0.040) | 0.636 (0.038) | 0.636 (0.416) | 0.630 (0.037) | 0.626 (0.038) |
Leukemia data | AUC of the test data by DLDA | ||||
t | t _{1,1} | s _{1,1} | s _{1,5} | s _{1,10} | |
10 genes | 0.981 (0.017) | 0.982 (0.017) | 0.986 (0.014) | 0.991 (0.012) | 0.990 (0.016) |
50 genes | 0.992 (0.014) | 0.992 (0.014) | 0.988 (0.013) | 0.994 (0.008) | 0.994 (0.008) |
100 genes | 0.992 (0.016) | 0.991 (0.017) | 0.989 (0.013) | 0.995 (0.006) | 0.995 (0.009) |
Gene ranking is an essential in biological investigations. In this study, we were motivated by the desire to identify robust and predictive biomarkers. Hetero genes may be informative for some patients, but uninformative in others. In this sense, hetero genes should be extracted from gene rankings if these predictive performance is eqaul to or less than that of homo genes.
Conclusions
The t-statistic confuses homo and hetero genes as shown in the simulation study. The ranking irreproducibility would be caused by such heterogeneity also in the real data analysis. In fact, even the signs of t-statistics of many genes mismatch in the real data. We present the sign-sum statistic for getting robust ranking. Robustness for heterogeneity of the sign-sum statistic is shown by the evaluation of the upper confidence limit. We can get more reproducible ranking by the sign-sum statistic for simulated data which assumes that there are heterogeneous factors, for the breast cancer data which is known as the hetero disease and the data which includes different disease statuses.
Availability of supporting data
The data sets supporting the results of this article are provided at the following database http://bioinformatics.nki.nl/data/van-t-Veer_Nature_2002/ and GEO under the accession number of GSE31568.
Abbreviations
AUC, area under curve; DLDA, diagonal linear discriminant analysis
Declarations
Acknowledgements
We thank the reviewers of our manuscript for careful reading and for giving beneficial suggestions. This work was supported by JSPS KAKENHI Grant Number 25280008.
Authors’ contributions
KO, OK and SE designed the methods of this article. KO carried out the simulation study and data analysis, and wrote the paper. All authors have read and approved the final manuscript.
Competing interests
The authors declare that they have no competing interests.
Open Access This article is distributed under the terms of the Creative Commons Attribution 4.0 International License(http://creativecommons.org/licenses/by/4.0/), which permits unrestricted use, distribution, and reproduction in any medium, provided you give appropriate credit to the original author(s) and the source, provide a link to the Creative Commons license, and indicate if changes were made. The Creative Commons Public Domain Dedication waiver(http://creativecommons.org/publicdomain/zero/1.0/) applies to the data made available in this article, unless otherwise stated.
Authors’ Affiliations
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