- Methodology
- Open Access
- Open Peer Review

# A novel joint analysis framework improves identification of differentially expressed genes in cross disease transcriptomic analysis

- Wenyi Qin
^{1}and - Hui Lu
^{1, 2, 3}Email authorView ORCID ID profile

**Received:**11 November 2017**Accepted:**29 January 2018**Published:**20 February 2018

## Abstract

### Motivation

Detecting differentially expressed (DE) genes between disease and normal control group is one of the most common analyses in genome-wide transcriptomic data. Since most studies don’t have a lot of samples, researchers have used meta-analysis to group different datasets for the same disease. Even then, in many cases the statistical power is still not enough. Taking into account the fact that many diseases share the same disease genes, it is desirable to design a statistical framework that can identify diseases’ common and specific DE genes simultaneously to improve the identification power.

### Results

We developed a novel empirical Bayes based mixture model to identify DE genes in specific study by leveraging the shared information across multiple different disease expression data sets. The effectiveness of joint analysis was demonstrated through comprehensive simulation studies and two real data applications. The simulation results showed that our method consistently outperformed single data set analysis and two other meta-analysis methods in identification power. In real data analysis, overall our method demonstrated better identification power in detecting DE genes and prioritized more disease related genes and disease related pathways than single data set analysis. Over 150% more disease related genes are identified by our method in application to Huntington’s disease. We expect that our method would provide researchers a new way of utilizing available data sets from different diseases when sample size of the focused disease is limited.

## Keywords

- Public data integration
- Cross disease transcriptome
- Gene expression
- Differentially expressed

## Introduction

High-throughput technology like microarray and next-generation sequencing (NGS) allows researchers measure thousands of gene or microRNA expression in one sample simultaneously. Detecting differentially expressed (DE) genes between disease and normal control group is one of the most common analyses in genome-wide transcriptomic data. Differentially expressed genes are potential disease-related genes and could be used for generating biological hypothesis of disease mechanism, developing potential clinical diagnosis tools and investigating potential drug targets. This approach has been successfully applied in many complex diseases like cancers [10, 13] and diabetes [5, 33].

With the cost of microarray and next generation sequencing technique decreasing and stabilization of the experiment protocol, there are now over 1,000,000+ samples deposited in public databases such as Gene Expression Ominus (GEO) [6]. With this huge amount of public data available, it is now possible for researchers to perform cross disease transcriptomics comparison analysis. For example, Borjabad [1] compared the transcriptomes of postmortem brain tissues among HIV-associated neurocognitive disorders, Alzheimer’s disease and multiple sclerosis and found a large number of overlapped DE genes, indicating the shared mechanism among these three diseases which might lead to a common therapeutic approach. Swindell [27] identified common and specific gene signature in psoriasis by comparing the DE genes in psoriasis transcriptome with other DE genes of similar skin diseases. The cross disease transcriptomic analysis has provided researchers with new opportunities of understanding of mechanisms of complex disease and discovery of new biomarkers.

Numerous cross-disease analyses have shown that similar diseases might share similar disease related genes. However, in these cross-disease comparison studies, they took a simple “disease-by-disease” approach: each disease was analyzed with traditional DE detection method like two-sample t-test or *limma* [25] separately, then the overlap of DE genes between diseases was examined. This approach falls short in its ability to jointly analyze data on all diseases to improve the identification power while simultaneously considering for difference among DE genes present in each disease. Because of the incomplete power, this simple approach might lead to difficulties in interpreting the result of whether a gene is commonly shared by all disease or specific to one disease. On the other hand, joint analysis methods developed in other fields of omics data analysis and have been proven a useful method to increase the identification power by borrowing information from other similar diseases [2, 3, 16, 30].

Meta-analysis approach is a popular data integrating statistical methods used to analyze multiple public datasets of same biological conditions [20]. They improve the identification power by detecting the weak yet consistent signals through all studies of the same purpose. However, they are not suitable for cross-disease transcriptomic analysis because they assume that a gene is either differentially expressed in all studies or non-differential in all studies [9, 21, 22] while ignoring the context-specific signals within each disease study.

Motivated by this, we propose an empirical Bayesian based mixture model which jointly analyzed multiple similar diseases to increase the identification power of common and disease-specific DE genes. The rest of paper is organized as follows. First, through a comprehensive simulation study, we compare our method with single data set analysis as well as two popular meta-analysis approaches with different underlying null hypothesis: minP, maxP [31, 32] and demonstrate that our method outperforms these methods in terms of identification power. Then we apply our method to two real cases by jointly analyzing six microarray studies of different cancers as well as Alzherimer’s disease (AD) and Huntington’s disease (HD) show that joint analysis identifies more DE genes than single data set analysis and these DE genes are enriched with disease-related genes and pathways.

## Methods

### Joint analysis framework formulation

Assume that there are N data sets. Each data set contains both disease and normal samples. Same G genes’ expressions are measured in each data set. In the proposed joint analysis framework, a differential test is first performed for each gene *g* (*g* = 1, …*G*) to obtain a differential test score within each data set *i* (*i* = 1, …, *N*). In this study, we choose to use two sample t-statistic: *t*_{
gi
} as the differential test score. We then transform *t*_{
gi
} into a Z-score: *Z*_{
gi
} according to McLachlan’s normal transformation [17] so that the Z-score distribution of non-DE genes will be approximately normal. These Z-scores will serve as the basis of inference of DE in the joint analysis framework. We assume a two-component mixture model for all genes’ Z scores within each data set *i* where each gene’s hidden DE status variable *D*_{
i
} is either DE (*D*_{
i
} =1) or non-DE (*D*_{
i
} =0). Then we assume two different conditional density distributions of Z-scores depending on a gene’s hidden status *D*_{
i
} in data set *i*: *f*(*Z*| *D*_{
i
}) where *D*_{
i
} =1 or 0. By doing so, we model the study-specific variation of Z-scores observed within each data set *i*.

*N*disease datasets, we want to compute the posterior probability to infer the DE status of gene

*g*in disease data set

*i*which could be written as:

*D*

_{ i }is determined, written as:

*f*(

*Z*|

*D*

_{ i }) distribution is different from data set to data set, so it needs to be estimated separately for each data set. Here we apply the method of local false discovery rate (local FDR) developed by Efron [7] to estimate this conditional distribution. We refer interested readers to Efron’s paper for the details of the method. Here we just briefly describe the estimation procedure. The local FDR is written as:

*f*(

*Z*

_{ gi }) =

*f*(

*Z*

_{ gi }|

*D*

_{ i }= 0) Pr(

*D*

_{ i }= 0) +

*f*(

*Z*

_{ gi }|

*D*

_{ i }= 1) Pr(

*D*

_{ i }= 1) and Pr(

*D*

_{ i }= 1) =1 − Pr(

*D*

_{ i }= 0).

In the localFDR approach, the marginal density *f*(*Z*_{
gi
}) is estimated through fitting z-scores of all genes to a cubic spline. The conditional density *f*(*Z*_{
gi
}| *D*_{
i
} = 0) is assumed to be a normal distribution. Its mean and variance as well as the quantity Pr(*D*_{
i
} = 0) are estimated through fitting the Z-scores in the central peak (around 0) by maximum likelihood estimation approach. This is a reasonable assumption because most Z-scores around the 0 should come from the non-DE distribution. Then through Eq. (3), we could also obtain the estimate of *f*(*Z*_{
gi
}| *D*_{
i
} = 1). All the estimation procedures described above are done through the *locfdr* package in R [8]. In this study, we also use Pr(*D*_{
i
} = 1| *Z*_{
gi
}) = 1− localFDR(*Z*_{
gi
}) computed by local FDR method as the inference result of single data set analysis for method comparison purpose.

*D*

_{1},

*D*

_{2}, …

*D*

_{ i }, …

*D*

_{ N }). The shared information between similar diseases is also modeled by this quantity. For example, we would expect that if one gene is DE in one disease, it is highly likely to be DE in another similar disease. In mathematics, we write this relation as Pr(

*D*

_{1}= 1,

*D*

_{2}= 1) = Pr(

*D*

_{2}= 1|

*D*

_{1}= 1) Pr(

*D*

_{1}= 1) and Pr(

*D*

_{1}= 1,

*D*

_{2}= 1) ≠ Pr(

*D*

_{1}= 1) Pr(

*D*

_{2}= 1). This prior probability could be estimated by using Expectation Maximization (EM) algorithm [4] to maximize the marginal log likelihood of all genes’ expression Z-scores in all data sets. EM algorithm steps could be summarized as follows:

- (1)
Initialize the prior probability: \( {\Pr}^{(0)}\left({D}_1,{D}_2,\dots {D}_i,\dots {D}_N\right)=\frac{1}{2^N} \).

- (2)
At iteration

*s*, compute the joint posterior probability of gene*g*given \( \overrightarrow{Z_g} \):

- (3)
Estimate the new prior probability at iteration s + 1 by averaging the joint probability calculated in step (2) over all genes: \( {\Pr}^{\left(s+1\right)}\left({D}_1,\dots, {D}_N\right)=\frac{1}{G}\sum \limits_{g=1}^G{\Pr}^{(s)}\left({D}_1,\dots, {D}_N\ |{Z}_{g1},\dots, {Z}_{gN}\right) \)

- (4)
Repeat step (2) and (3) until convergence.

The proposed joint analysis framework is implemented under R statistical programming language.

### Simulation studies

In real world, researchers often have limited samples for a specific disease and do not have other public data sets of the same disease while public data set of other similar diseases exists. We design a simulation study which mimics the real situation to test and compare the performance of our method with others. Our simulation study models this situation by generating different number of studies with similar but slightly different DE gene configuration in each disease.

To be more specific, the simulation is set to have *N* studies with 15 disease and 15 control samples within each study. Each study here could be considered as a similar disease. There is a total of 10,000 genes expression value measured in each sample. We first need to determine the hidden DE status variable value for each gene in each study. We define *shared percentage* between study *i* and study *j* as the conditional probability of being DE in study *j* if the hidden DE status variable in study *i* is true, i.e. Pr(D_{
j
} = 1 ∣ D_{
i
} =1). We further define “similarity” as the average shared percentage of DE genes between two studies, i.e. \( \frac{1}{2}\left(\Pr \left({\mathrm{D}}_j=1|{\mathrm{D}}_i=1\right)+\Pr \left({\mathrm{D}}_i=1|{\mathrm{D}}_j=1\right)\right)=\frac{1}{2}\left(\frac{\Pr \left({D}_i=1,{D}_j=1\right)}{\Pr \left({\mathrm{D}}_i=1\right)}+\frac{\Pr \left({D}_i=1,{D}_j=1\right)}{\Pr \left({\mathrm{D}}_j=1\right)}\right) \). We also assume there is around 10% of DE genes in each study i.e. Pr(D_{
i
} =1) = 0.1. We finally define two diseases are “similar” if the similarity value between two diseases is higher than the expected similarity (i.e. Pr(*D*_{
i
} = 1, *D*_{
j
} = 1) = Pr(D_{
i
} = 1) Pr(D_{
j
} = 1)). Therefore, once the DE status variable value of a gene in the reference study is known, we could generate DE status variables of this gene for all other studies. In this simulation, we assume study 1 is the reference study, the hidden DE status configuration of other studies is then generated for each gene based on the DE status variable in study. After the hidden DE status variable is determined, we generate the normally distributed expression value based on the DE status of each gene. The variance σ^{2}_{
gd
} of each gene *g* is assumed to be the same in each study *d* and is sampled from an inverse chi-square distribution with degrees of freedom 4 and scale parameter 0.02. We then generate gene ’s expression for every sample from N(0, σ^{2}_{
gd
}). If D_{
i
} =1, we sample a μ_{
gd
} from N(0, *w*_{
gd
} ∗ σ^{2}_{
gd
}) where *w*_{
gd
} = 4 here and add it to the expression value of disease samples. By using this simulation setup, we mimic the real case when the sample size of target disease is small but studies of similar diseases exist in public database.

We also design another simulation study by fixing the hidden DE status of each gene before generating the expression value. In this simulation study, we assume that there are *N* = 2 data sets, with same number of genes and samples setup described above. In the first data set, the first 1000 genes are assumed to be DE. In the second data set, we assume that first *X* genes are DE and for the rest of *1000-X* genes, we ensure that they will not overlap with any DE genes in data set 1. Once the DE status of all genes are set in two data sets, gene expression values are generated with the same procedure described above. By setting so, the true prior probability is a fixed value. For example, if X = 600, then the prior probability will be Pr(D_{1} = 1, D_{2} =1) =0.06, Pr(D_{1} = 0, D_{2} =1) =0.04, Pr(D_{1} = 1, D_{2} =0) =0.04 and Pr(D_{1} = 0, D_{2} =0) = 0.86 respectively. By using this simulation setup, we are able to compare the estimated prior probability generated from joint analysis with the true value.

### Meta-analysis methods

Two popular meta-analysis approaches are compared with our joint analysis method: minP and maxP [28]. These two methods represent two different underlying hypothesis used in meta-analysis methods: the first hypothesis tests if one gene is DE in at least one or more data sets or not; the second hypothesis detects if one gene is DE in all studies or not [28]. Briefly speaking, The maxP method takes maximum of *p*-value from each study as test statistics: *S*_{
g
}^{
maxP
} = *argmax*(*p*_{
gk
}). *S*_{
g
}^{
maxP
} follows a beta distribution with degrees of freedom α = *N* and β = 1 under null hypothesis. The maxP method targets the DE genes with small *p*-values in all studies. The minP method takes the minimum p-value among the K studies as the test statistic: *S*_{
g
}^{
minP
} = *argmin*(*p*_{
gk
}). It follows a beta distribution with degrees of freedom α = 1 and β = N under the null hypothesis. This method detects a DE gene whenever a small p-value exists in any one of all studies. All meta-analysis methods used in this paper are implemented through using *metaRawdata()* function in *metaDE* R package [31, 32]. Two sample t-test is used as summary statistic for each individual study and parametric assumption is used to obtain the p-value of the statistic.

### Real data application

#### Cancer data sets

Summaries of six different cancer data sets used in this study

#### Alzheimer’s disease and Huntington’s disease data sets

Narayanan et al. conducted a co-expression network analysis between Alzheimer’s disease (AD) and Huntington’s disease (HD) using the prefrontal cortex region of postmortem brain samples consisting of 310 AD patients, 157 HD patients and 157 controls [19]. The microarray expression data is deposited at GEO (GEO Accession no: GSE33000) and downloaded. A linear model is then fit to each gene with gender, age and hidden batch variables estimated with *sva* R package [15] as covariates to correct for confounding factors. The t-statistic of disease effect of each gene is then extracted. Single data set analysis and joint analysis are then applied to the t statistics and DE results are obtained for each disease. A total of 39,280 probes (some probes will represent the same gene) are measured and will be used in this study.

## Results

### Formulation of proposed joint analysis framework

*g*in each data set

*i*, then the differential test statistics is transformed into a Z-score:

*Z*

_{ gi }. Then within each data set, estimate

*f*(

*Z*|

*D*

_{ i }) distribution with localFDR approach. After the conditional density value is obtained for each

*Z*

_{ gi }, the prior probability Pr(

*D*

_{1},

*D*

_{2}, …

*D*

_{ i }, …

*D*

_{ N }) is estimated through EM algorithm. Finally compute posterior probability defined in Eq. (2) in Methods section for each gene in each data set and genes are ranked based on this quantity. A gene would be called DE if the posterior probability is higher than some pre-defined threshold.

### Simulation studies

#### Comparison between joint analysis and single data set analysis

We begin by comparing the identification power between single data set analysis and joint analysis using simulation studies. Different simulated disease data sets are generated by varying the number of data sets and shared percentage among diseases as described in “Methods” section. The number of data sets is set for *N* = 1, 2, 4, 6 and the shared percentage between study 1 and other studies is set to Pr(D_{
j
} = 1 ∣ D_{1} =1) = 0,0.1, 0.6, 0.7, 0.8, 0.9, 1. Every parameter combination is repeated for 100 times. In each run, we set a specified posterior probability cutoff in data set 1 which is considered as the disease data set of interest and report the average sensitivity as well as average false discovery rate (FDR) in study 1 as the result. The cutoff is set to 0.95.

#### Comparison with other meta-analysis approaches

*N*= 2, 4, 6 and the shared percentage between study 1 and other studies is set to Pr (D

_{j}=1|D

_{1}= 1) = 0.6, 0.8, 1. The results are shown in Fig. 3. When the shared percentage is set to 0.6, the joint analysis consistently outperforms all other methods by identifying more true DE genes in top 1000 genes except in the rank range of 900 to 1000 when

*N*= 2 minP and joint analysis have similar performance. When the shared percentage value increases to 0.8, joint analysis outforms other methods in top ranking below 800, minP method performs better in the rank range of top 800–1000 genes. When the shared percentage is 1, which means all measurement are based on same disease, minP has better performance. Overall, when the data sets are based on similar but different diseases, especially when more diseases are included, our joint analysis outperformed other methods.

#### Evaluation of estimated prior probability

*X*value was set to 600, 700, 800, 900 and 1000. Each parameter setup was repeated 10 times and the results were summarized in Table 2. By comparing the estimated value with true value, we observed that the joint analysis framework will underestimate the shared percentage of genes but had an increasing trend when the shared number of genes increases. Further, when the number of data sets increased to 4 and 6, we compared the similarity estimate obtained from joint analysis with true similarity value as defined in the “Methods” section between data set 1 and data set 2. The simulation setup was the same as that in the sensitivity and FDR comparison and the results of comparison were similar (Additional file 2: Table S2). The main reason for the observed conservative estimate of shared gene pairs might be that the local false discovery method implemented in the joint analysis framework tends to be conservative by classifying most genes at the boundary between the null distribution and alternative distribution to the null distribution so that the shared gene pairs at the boundary might be difficult to be correctly classified. This problem could be alleviated by employing parametric distribution setup for the joint analysis framework but the current non-parametric framework is more general and could be used in more situations. Nevertheless, the accurate estimation of increasing trend of shared gene pairs could help the joint analysis to put correct priors among diseases to infer DE status of a gene.

Comparison of estimated prior probability with true ratio in the simulation study

DE status | X | |||||||||
---|---|---|---|---|---|---|---|---|---|---|

600 | 700 | 800 | 900 | 1000 | ||||||

Estimate | Truth | Estimate | Truth | Estimate | Truth | Estimate | Truth | Estimate | Truth | |

(0,0) | 0.8857 (0.006) | 0.86 | 0.8914 (0.006) | 0.87 | 0.8982 (0.006) | 0.88 | 0.9054 (0.005) | 0.89 | 0.9068 (0.003) | 0.9 |

(0,1) | 0.041 (0.005) | 0.04 | 0.0356 (0.005) | 0.03 | 0.0284 (0.004) | 0.02 | 0.0218 (0.003) | 0.01 | 0.018 (0.003) | 0 |

(1,0) | 0.042 (0.008) | 0.04 | 0.0365 (0.008) | 0.03 | 0.0316 (0.006) | 0.02 | 0.0258 (0.003) | 0.01 | 0.02 (0.004) | 0 |

(1,1) | 0.031 (0.003) | 0.06 | 0.0364 (0.002) | 0.07 | 0.0417 (0.004) | 0.08 | 0.0469 (0.005) | 0.09 | 0.0546 (0.004) | 0.1 |

#### Influence of sample size of a similar disease

### Real data application: six different cancers

*N*= 6 and increased shared percentage parameter setup with cutoff = 0.95. The simulation study suggested that the joint analysis could identify most of genes which are identified by single data set analysis (Additional file 4: Table S4). The comparison results of cancer data sets were thus consistent with those in simulation studies and demonstrated that our proposed joint analysis framework could identify most of genes that are also identified by single data set analysis with improved identification power.

To further validate the biological relatedness of identified DE genes, we also checked if the DE gene lists are enriched with cancer-related genes by comparing the DE gene lists with a 743 known cancer-related gene lists compiled by Nagaraj [18]. The results in Fig. 5b showed that the joint analysis identifies more cancer-related genes than single data set analysis and hypergeometric test shows that the newly identified DE genes are enriched with cancer-related genes in bladder, colorectal, lung and NSCLC data sets respectively while there is no enrichment seen in the results of single data set analysis using the same cutoff.

Pair-wise similarity estimated among cancers

Bladder | Colorectal | NSCLC | AML | Lung | Pancreatic | |
---|---|---|---|---|---|---|

Bladder | 1 | 0.521 | 0.525 | 0.219 | 0.435 | 0.132 |

Colorectal | 0.521 | 1 | 0.511 | 0.168 | 0.486 | 0.117 |

NSCLC | 0.525 | 0.511 | 1 | 0.251 | 0.906 | 0.148 |

AML | 0.219 | 0.168 | 0.251 | 1 | 0.197 | 0.085 |

Lung | 0.435 | 0.486 | 0.906 | 0.197 | 1 | 0.124 |

Pancreatic | 0.132 | 0.117 | 0.148 | 0.085 | 0.124 | 1 |

### Real data application: Alzheimer’s disease and Huntington’s disease

*p*-values. The number of DE genes identified along the pathway, the raw enrichment

*p*-value of the pathway and Bonferroni’s corrected

*p*-value are reported in Table 6. Table 6 showed that the enriched KEGG pathways obtained by single analysis and joint analysis have a large overlap. The joint analysis prioritized three similar neurodegenerative disease related pathways and their corresponding biological process: oxidative phosphorylation over single data set analysis by identifying more DE genes along those pathways shared by these diseases. Metabolic pathways are found to be differentially expressed in several neurodegenerative disorders such as in schizophrenia [20] and identified as more enriched in joint analysis. It is worth noting that synaptic vesicle cycle pathway, which is closely related to neurotransmitter release and neurodegenerative disorders [34], is exclusively identified by joint analysis.

Number of genes in KEGG pathway of AD and HD among top ranked genes in each neurodegenerative disorder

Alzheimer’s Disease | Huntington’s disease | |||
---|---|---|---|---|

Top Rank | Single | Joint | Single | Joint |

< 250 | 2 | 4 | 5 | 6 |

< 500 | 9 | 12 | 10 | 16 |

< 750 | 19 | 21 | 14 | 24 |

< 1000 | 29 | 30 | 19 | 32 |

Posterior probability and rank comparison of 15 HD-related genes exclusively identified by joint analysis among top 1000 genes

Gene Symbol | Single P | Joint P | Single Rank | Joint Rank |
---|---|---|---|---|

ATP5B | 0.722686833 | 0.938701726 | 1475 | 652 |

ATP5F1 | 0.712111394 | 0.933163998 | 1690 | 954 |

ATP5G1 | 0.724132275 | 0.939691574 | 1443 | 594 |

ATP5J | 0.71465263 | 0.935545911 | 1639 | 827 |

CLTA | 0.704691436 | 0.933779808 | 1833 | 918 |

COX4I1 | 0.732382138 | 0.938415838 | 1208 | 673 |

NDUFA7 | 0.718914525 | 0.938173615 | 1553 | 683 |

NDUFA9 | 0.738565931 | 0.942103957 | 1037 | 460 |

NDUFB5 | 0.731956544 | 0.940700886 | 1218 | 546 |

NDUFB6 | 0.709427659 | 0.932649277 | 1741 | 972 |

POLR2K | 0.705645646 | 0.935145538 | 1806 | 850 |

SLC25A5 | 0.737126871 | 0.934489089 | 1089 | 884 |

UQCRC1 | 0.727082935 | 0.932648306 | 1373 | 973 |

UQCRH | 0.723833736 | 0.934133476 | 1448 | 902 |

VDAC2 | 0.738319827 | 0.944168188 | 1045 | 327 |

Top 10 KEGG pathway enrichment results comparison between (A) single data set analysis and (B) joint analysis in Huntington’s disease

Term | Count | Enrichment Pvalue | Bonferroni corrected P |
---|---|---|---|

(A) Single | |||

hsa03050:Proteasome | 12 | 5.14E-07 | 1.21E-04 |

hsa05010:Alzheimer’s disease | 20 | 1.95E-05 | 0.004579209 |

hsa05012:Parkinson’s disease | 18 | 2.58E-05 | 0.006034046 |

hsa05016:Huntington’s disease | 19 | 3.66E-04 | 0.08242395 |

hsa05033:Nicotine addiction | 7 | 0.003777846 | 0.589128604 |

hsa00190:Oxidative phosphorylation | 13 | 0.004721495 | 0.671158323 |

hsa04932:Non-alcoholic fatty liver disease (NAFLD) | 14 | 0.00497103 | 0.689975889 |

hsa05169:Epstein-Barr virus infection | 15 | 0.013743654 | 0.9613094 |

hsa04723:Retrograde endocannabinoid signaling | 10 | 0.01471692 | 0.969321024 |

hsa04728:Dopaminergic synapse | 11 | 0.02422846 | 0.996860832 |

(B) Joint | |||

hsa05012:Parkinson’s disease | 29 | 3.18E-13 | 7.59E-11 |

hsa00190:Oxidative phosphorylation | 27 | 2.90E-12 | 6.92E-10 |

hsa05016:Huntington’s disease | 32 | 4.34E-12 | 1.04E-09 |

hsa05010:Alzheimer’s disease | 29 | 2.33E-11 | 5.57E-09 |

hsa04932:Non-alcoholic fatty liver disease (NAFLD) | 22 | 2.14E-07 | 5.12E-05 |

hsa03050:Proteasome | 10 | 3.21E-05 | 0.007653523 |

hsa01100:Metabolic pathways | 73 | 5.47E-05 | 0.012999463 |

hsa05169:Epstein-Barr virus infection | 20 | 1.02E-04 | 0.024158392 |

hsa04721:Synaptic vesicle cycle | 10 | 5.70E-04 | 0.127417187 |

hsa01200:Carbon metabolism | 13 | 0.001156092 | 0.241540496 |

## Conclusion and discussion

In this paper, we present a novel statistical framework which aims at addressing a problem often met by biological researchers: when only a limited number of sample for a specific disease is available, the identification power could be improved by jointly analyzing multiple similar disease data sets because DE genes might be shared among similar diseases. By implementing a two-component mixture model, we demonstrate the framework could improve the identification power through comprehensive simulation studies and two real data applications. The joint analysis outperforms single data set analysis in both identification power and biological interpretation.

The prior probability is the most essential quantity in the proposed joint analysis framework and has a large impact on the performance of the method because similarity between diseases are directly determined by this quantity. This has been demonstrated through both simulation study and real data application. In simulation studies, we observed that when jointly analyzed with diseases with higher similarity, which was realized by adjusting prior probability value among diseases, the target data set gained more statistical power than less similar diseases. In real data application, more DE genes were identified among similar cancers than dissimilar ones where similarity among cancers were computed through estimated prior probability. In short, prior probabilities among different diseases could determine if the proposed joint analysis framework would be effective or not.

There would be several improvements for the proposed joint framework in the future. The first issue to be addressed is how to jointly analyze more disease data sets. As mentioned by one reviewer, the estimation of the prior probability in the proposed framework here is computationally intensive when the number of diseases to be jointly analyzed is large (~2^{N}, where N is the total number of diseases). The estimation of prior probability would become infeasible when the number reaches 20 or more. Some potential solution to this problem has been proposed in a recent paper [14]. The basic idea is to assume special structures about the prior probability such that the number of prior probability to be estimated could be significantly reduced, thus incorporating more disease data sets becomes available. Another improvement would be to design a disease similarity test so that researchers could determine if two diseases are similar enough to be jointly analyzed. A similar idea has been proposed by Chung et al. [3] where a likelihood test was designed to evaluate if two diseases contain similar SNPs. Finally, next generation sequencing support is expected to be added to current framework such that microarray and sequencing data could be analyzed simultaneously.

## Declarations

### Acknowledgements

We appreciate Dr. Hongyu Zhao’s insightful and helpful discussions in joint analysis of omics data, and Mr. Xujun Wang’s help in manuscript preparation.

### Funding

The whole project including design of the study and collection, analysis, and interpretation of data and in writing the manuscript is supported by University of Illinois at Chicago department of Bioengineering and Science and Technology Commission of Shanghai Municipality (STCSM) grant 17DZ 22512000.

### Availability of data and materials

The joint analysis R code is available at https://github.com/qintian0321/Joint-analysis-DE-gene.

### Authors’ contributions

WQ and HL conceived and designed the study. WQ analyzed data and interpreted the results, WQ and HL wrote the manuscript. Both authors read and approved the final manuscript.

### Ethics approval and consent to participate

Not applicable.

### Consent for publication

Not applicable

### Competing interests

The authors declare that they have no competing interests.

### Publisher’s Note

Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.

**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

## References

- Borjabad A, Volsky DJ. Common transcriptional signatures in brain tissue from patients with HIV-associated neurocognitive disorders, Alzheimer’s disease, and Multiple Sclerosis. J Neuroimmune Pharmacol. 2012;7:914–26.View ArticlePubMedPubMed CentralGoogle Scholar
- Chen X, Slack FJ, Zhao H. Joint analysis of expression profiles from multiple cancers improves the identification of microRNA–gene interactions. Bioinformatics. 2013;29(17):2137–45.View ArticlePubMedPubMed CentralGoogle Scholar
- Chung D, Yang C, Li C, Gelernter J, Zhao H. GPA: a statistical approach to prioritizing GWAS results by integrating pleiotropy and annotation. PLoS Genet. 2014;10(11):e1004787.View ArticlePubMedPubMed CentralGoogle Scholar
- Dempster AP, Laird NM, Rubin DB. Maximum likelihood from incomplete data via the EM algorithm. J R Stat Soc Ser B Methodol. 1977;39(1):1–38.Google Scholar
- Eckenrode SE, et al. Gene expression profiles define a key checkpoint for type 1 diabetes in NOD mice. Diabetes. 2004;53:366–75.View ArticlePubMedGoogle Scholar
- Edgar R, Domrachev M, Lash AE. Gene expression omnibus: NCBI gene expression and hybridization array data repository. Nucleic Acids Res. 2002;30:207–10.View ArticlePubMedPubMed CentralGoogle Scholar
- Efron B. Large-scale simultaneous hypothesis testing: the choice of a null hypothesis. J Am Stat Assoc. 2004;99:96–104.View ArticleGoogle Scholar
- Efron, Bradley, Brit Turnbull, and Balasubramanian Narasimhan. (2011) locfdr: computes local false discovery rates. R package.Google Scholar
- Hong F, et al. RankProd: a bioconductor package for detecting differentially expressed genes in meta-analysis. Bioinformatics. 2006;22:2825–7.View ArticlePubMedGoogle Scholar
- Hou J, et al. Gene expression-based classification of non-small cell lung carcinomas and survival prediction. PLoS One. 2010;5:e10312.View ArticlePubMedPubMed CentralGoogle Scholar
- Huang DW, Sherman BT, Lempicki RA. Systematic and integrative analysis of large gene lists using DAVID bioinformatics resources. Nat Protoc. 2009;4(1):44–57.View ArticleGoogle Scholar
- Ishikawa M, et al. Experimental trial for diagnosis of pancreatic ductal carcinoma based on gene expression profiles of pancreatic ductal cells. Cancer Sci. 2005;96:387–93.View ArticlePubMedGoogle Scholar
- Kim WJ, et al. Predictive value of progression-related gene classifier in primary non-muscle invasive bladder cancer. Mol Cancer. 2010;9:3.View ArticlePubMedPubMed CentralGoogle Scholar
- Lai Y, Zhang F, Nayak TK, Modarres R, Lee NH, McCaffrey TA. An efficient concordant integrative analysis of multiple large-scale two-sample expression data sets. Bioinformatics. 2017;33:3852.View ArticlePubMedGoogle Scholar
- Leek JT, Johnson WE, Parker HS, Jaffe AE, Storey JD. The sva package for removing batch effects and other unwanted variation in high-throughput experiments. Bioinformatics. 2012;28(6):882–3.View ArticlePubMedPubMed CentralGoogle Scholar
- Lin Z, Wang T, Yang C, Zhao H. On joint estimation of Gaussian graphical models for spatial and temporal data. Biometrics. 2017;73:769.View ArticlePubMedPubMed CentralGoogle Scholar
- McLachlan GJ, Bean RW, Jones LB. A simple implementation of a normal mixture approach to differential gene expression in multiclass microarrays. Bioinformatics. 2006;22:1608–15.View ArticlePubMedGoogle Scholar
- Nagaraj SH, Reverter A. A Boolean-based systems biology approach to predict novel genes associated with cancer: application to colorectal cancer. BMC Syst Biol. 2011;5:35.View ArticlePubMedPubMed CentralGoogle Scholar
- Narayanan M, Huynh JL, Wang K, Yang X, Yoo S, McElwee J, et al. Common dysregulation network in the human prefrontal cortex underlies two neurodegenerative diseases. Mol Syst Biol. 2014;10(7):743.View ArticlePubMedPubMed CentralGoogle Scholar
- Qin W, et al. Meta-analysis of sex differences in gene expression in schizophrenia. BMC Syst Biol. 2016;10(Suppl 1):9.View ArticlePubMedPubMed CentralGoogle Scholar
- Rhodes DR, et al. Large-scale meta-analysis of cancer microarray data identifies common transcriptional profiles of neoplastic transformation and progression. Proc Natl Acad Sci U S A. 2004;101:9309–14.View ArticlePubMedPubMed CentralGoogle Scholar
- Ruan L, Yuan M. An empirical Bayes’ approach to joint analysis of multiple microarray gene expression studies. Biometrics. 2011;67:1617–26.View ArticlePubMedGoogle Scholar
- Selamat SA, et al. Genome-scale analysis of DNA methylation in lung adenocarcinoma and integration with mRNA expression. Genome Res. 2012;22:1197–211.View ArticlePubMedPubMed CentralGoogle Scholar
- Sheffer M, et al. Association of survival and disease progression with chromosomal instability: a genomic exploration of colorectal cancer. Proc Natl Acad Sci U S A. 2009;106:7131–6.View ArticlePubMedPubMed CentralGoogle Scholar
- Smyth GK. Limma: linear models for microarray data. In: Bioinformatics and computational biology solutions using R and Bioconductor. New York: Springer; 2005. p. 397–420.View ArticleGoogle Scholar
- Stirewalt DL, et al. Identification of genes with abnormal expression changes in acute myeloid leukemia. Genes Chromosomes Cancer. 2008;47:8–20.View ArticlePubMedGoogle Scholar
- Swindell WR, et al. Cross-disease transcriptomics: unique IL-17A signaling in psoriasis lesions and an autoimmune PBMC signature. J Invest Dermatol. 2016;136:1820–30.View ArticlePubMedPubMed CentralGoogle Scholar
- Tseng GC, Ghosh D, Feingold E. Comprehensive literature review and statistical considerations for microarray meta-analysis. Nucleic Acids Res. 2012;40:3785–99.View ArticlePubMedPubMed CentralGoogle Scholar
- Wang J, Liu Q, Shyr Y. Dysregulated transcription across diverse cancer types reveals the importance of RNA-binding protein in carcinogenesis. BMC Genomics. 2015;16(Suppl 7):S5.View ArticlePubMedPubMed CentralGoogle Scholar
- Wang T, Chen M, Zhao H. Estimating DNA methylation levels by joint modeling of multiple methylation profiles from microarray data. Biometrics. 2016;72(2):354–63.View ArticlePubMedGoogle Scholar
- Wang X, et al. An R package suite for microarray meta-analysis in quality control, differentially expressed gene analysis and pathway enrichment detection. Bioinformatics. 2012a;28:2534–6.View ArticlePubMedPubMed CentralGoogle Scholar
- Wang X, et al. Detecting disease-associated genes with confounding variable adjustment and the impact on genomic meta-analysis: with application to major depressive disorder. BMC Bioinf. 2012b;13:52.View ArticleGoogle Scholar
- Wilson KH, et al. Microarray analysis of gene expression in the kidneys of new- and post-onset diabetic NOD mice. Diabetes. 2003;52:2151–9.View ArticlePubMedGoogle Scholar
- Zoltowska KM, Maesako M, Lushnikova I, Takeda S, Keller LJ, Skibo G, et al. Dynamic presenilin 1 and synaptotagmin 1 interaction modulates exocytosis and amyloid β production. Mol Neurodegener. 2017;12(1):15.View ArticlePubMedPubMed CentralGoogle Scholar