Operon information improves gene expression estimation for cDNA microarrays
© Xiao et al; licensee BioMed Central Ltd. 2006
Received: 30 September 2005
Accepted: 21 April 2006
Published: 21 April 2006
In prokaryotic genomes, genes are organized in operons, and the genes within an operon tend to have similar levels of expression. Because of co-transcription of genes within an operon, borrowing information from other genes within the same operon can improve the estimation of relative transcript levels; the estimation of relative levels of transcript abundances is one of the most challenging tasks in experimental genomics due to the high noise level in microarray data. Therefore, techniques that can improve such estimations, and moreover are based on sound biological premises, are expected to benefit the field of microarray data analysis
In this paper, we propose a hierarchical Bayesian model, which relies on borrowing information from other genes within the same operon, to improve the estimation of gene expression levels and, hence, the detection of differentially expressed genes. The simulation studies and the analysis of experiential data demonstrated that the proposed method outperformed other techniques that are routinely used to estimate transcript levels and detect differentially expressed genes, including the sample mean and SAM t statistics. The improvement became more significant as the noise level in microarray data increases.
By borrowing information about transcriptional activity of genes within classified operons, we improved the estimation of gene expression levels and the detection of differentially expressed genes.
Genome-wide monitoring of transcription by means of DNA microarrays is used to infer transcriptional and regulatory networks in living organisms. In most of microarray experiments, transcript levels of thousands of genes are measured with a relatively small number of replications, so the estimates of true expression levels from microarray data may be poor, mostly due to a small sample size. To address this problem, several statistical methods have been proposed to borrow information from other genes to improve detection of the differentially expressed ones [1–9]. The main idea is to borrow information from other genes to estimate either the distributions of genes's expression levels or the distribution of error terms. The underlying assumption is that it should be possible to improve the estimates of expression levels of genes by borrowing information about transcriptional activity across the sets of genes that are biologically, or physically, related. In some cases, the expression levels may significantly vary across the genes, then borrowing information from unrelated genes may not improve, or even worsen, the estimates of gene expression levels . However, if, based on biological knowledge, we can expect that some genes are more likely to express at similar levels (i.e. co-express), then we can improve the inference by using information about the activity of those genes.
An operon  is a set of linearly juxtaposed genes transcribed as a single mRNA; operons are commonly found in prokaryotic genomes such as Escherichia coli. Transcription of operons of E. coli has been examined, and operons have been predicted in many studies [12–19], which provides background information about the E. coli regulatory network. The genes within the same operon usually have similar expression levels, hence show some local structure in expression profiles , and this fact has been successfully used in operon prediction [21, 22].
Based on the existing information about the structure of operons, we propose a hierarchical Bayesian model which improves gene expression estimation by borrowing information from genes within the same operon. Most existing methods for detecting differently expressed genes [1–9] borrow information from other genes in the whole genome, while our proposed method only borrows information from other genes within the same operon, which has sound biological basis. Wren et al  have proposed a simulated annealing approach to adjust gene expression data by using existing microarray measurements obtained on the same organism, which effectively reduced the noise and made it possible to compare different microarray experiments. But their method relies on reference microarray experiments that cover the dynamic range of transcript abundances for most of the genes, which may be difficult to select or unavailable. Instead of using existing microarray measurements, we use existing information about the operons' structure to reduce the noise in microarray data.
A more accurate estimation of transcript abundances of individual genes will improve our ability to evaluate transcriptional activity on a genome-wide scale, and hence facilitate the exploration of gene regulatory networks. Our proposed method provides a better way to estimate relative transcript levels, which is critical for distinguishing differentially expressed (DE) genes from equally expressed (EE) genes. Herein, we refer to the logarithm of a ratio of the fluorescent intensities of the test and control samples as the observed gene expression level. The genes with estimated expression levels significantly different from zero are identified as DE genes, otherwise as EE genes.
Using more than 200 microarray experiments, we obtained the evidence of co-transcription of genes within E. coli operons on a genome-wide scale. We applied the proposed method to three simulated and one experimentally obtained data sets. The simulation studies and the real data application demonstrated that the proposed method performed better than the sample mean and the SAM t statistics in estimating the gene expression levels as well as in detecting differentially expressed genes. The improvement became more significant as the noise level in microarray data increased.
Mean squared errors for different methods
Application to E. coli data
Top 44 genes and their estimated relative transcription levels
We examined some genes and operons in more detail, to demonstrate the advantages of borrowing information from within an operon. For example, an operon argT-hisJQMP contains 5 genes (argT, hisJ, hisM, hisP, hisQ) and is not expected to be differentially expressed under the examined experimental condition, according to our biological knowledge . But from the microarray data, the mean expression level (an average log ratio) of the gene argT was -2.73, which ranked 15th among all the expression levels in the E. coli genome. This "high expression" of the argT could possibly be caused by random noise in microarray measurements and/or in biological samples, since the mean expression levels of other genes within the same operon were close to 0 (those for genes hisP, hisM, hisQ, hisJ were 0.00, -0.05, 0.03, and 0.05, respectively). However, accounting for expression levels of other genes within the same operon lowered the estimate of the expression level (posterior mean of μ i of the argT to -0.02, rank of 1456, indicating that this gene was not differentially expressed. Analysis of another operon, fliDST, illustrates a complimentary case. Transcription of the fliDST operon (containing 3 genes, fliD, fliS, fliT) is known to be controlled by the FlhDC , and thus, under the experimental condition, differential expression of genes in that operon would be expected. While the expression level of the fliT, estimated by the sample mean at -0.90, ranked only 65th, the estimated expression level of the gene after borrowing information from two other genes in the operon was -3.25, which ranked 19th.
In this paper, we proposed and applied a hierarchical Bayesian model, to estimate relative gene expression levels and detect differentially expressed genes by borrowing expression information within operons. The performance of the proposed method was compared with that of the sample mean and SAM t statistics. Through the simulation studies, we showed that the proposed method outperformed the sample mean and the SAM t statistics in estimating gene expression levels and detecting DE genes. The proposed method was used to analyze differential expression in an E. coli mutant with a defect in transcription of motility/chemotaxis genes, giving results more consistent with the existing biological knowledge than those obtained by using the other statistics.
A major advantage of the proposed approach is in borrowing expression information from other genes within the same operon. The approach is developed within a statistically sound Bayesian model and it offers necessary flexibility with respect to the amount of information that needs to be borrowed from other genes. By borrowing information we can obtain stabilized estimates of expression levels from rather noisy microarray data. As a result, the estimates of transcript levels within the same operon become more similar to each other, more so than without borrowing information; this is consistent with a biological fact that genes within the same operon are transcribed as a single mRNA molecule. With the proposed method, the estimated expression levels of genes in "differentially expressed" operons are consistently high, and more importantly, transcript abundances of genes in "equally expressed" operons are stabilized towards zero. In the experimentally obtained microarray data, the within operon variation was smaller than the variation among the replicate data points for the same genes, indicating that the expression levels of the genes within an operon were very similar.
In our model, the ratio of parameters τ2 and determines how much information comes from the observed expression of a gene and how much comes from the average expression of an operon, when estimating the expression level of an individual gene within an operon. A smaller τ, as compared to , puts more weight on the average expression of an operon. Here we assumed the same τ value for all operons, implying that all operons had similar within-the-operon variability. However, this assumption might not be realistic. In some operons and physiological conditions, the genes might express very similarly, but in others, especially under the control of internal promoters, transcription of individual genes may be more heterogeneous. In the future, we will investigate the effect of an operon-specific τ. Operonal organization of genes is common in prokaryotes and also present in some eukaryotic organisms [28–30], and the proposed method can be extended to biological systems where the operonal structure is unknown. Many biological studies have demonstrated that co-expressed genes tend to cluster on the chromosome. Although the nature of this phenomenon is not quite understood, a positional clustering of co-expressed genes can be found in many eukaryotes including yeast [31, 32], worm , fly [34, 35], mouse , and human [37–39]. These findings indicate that the genes are likely to co-transcribe with their chromosomal neighbors. In those cases, instead of borrowing information from genes in the same operon, we can borrow information from gene neighbors on the chromosome. Another extension of our method would involve incorporation of gene annotation information into the analysis of expression data. The approach would be very similar to the one described in this paper: based on biological knowledge, the genes belonging to the same functional group are more likely to be co-expressed, so we can use a hierarchical model to borrow information from and for the genes within the same functional group to improve the estimates of gene expression levels.
The information about operon structure leads to a better estimation of gene expression levels. Using simulated and experimental data sets, we have demonstrated that the proposed method performs better than the sample mean and the SAM t statistics in estimating the relative levels of transcript abundances and detecting differentially expressed genes.
RegulonDB database and E. coli. microarray data
RegulonDB  is a database containing information about known operons in E. coli. According to the RegulonDB annotations, 1486 genes (about one third of all genes predicted in the E. coli genome) are organized in 600 operons.
The E. coli data set contains results of 217 microarrays collected in 53 different experimental conditions. The fluorescent intensities of the test and control samples were measured, and the average log ratio of the intensities for each gene under the same condition was used here to represent an observed gene expression level under that condition .
An E. coli motility expression data set ( series accession number: GPL2101) was obtained in a direct pair-wise comparison between a knock-out mutant of the flhDC, a master regulator of the motility/chemotaxis regulon , and its isogenic wild type strain. Total RNA samples of a mutant E. coli (test samples) and an isogenic wild type E. coli (control samples) were labeled with red (Cy5) and green (Cy3) fluorophors. The intensities from the red and the green channels were normalized by the lowess method . There were 4281 genes (G = 4281) with four replicates for each gene (n = 4). Let Yi,jbe defined as the log ratio of the intensities between the test and control samples for gene i on array j; that is,
We propose a hierarchical Bayesian model,
where, Yi,jis the log ratio of gene i in replicate j, μ i and σ i are the true expression level and the standard deviation, respectively. In our method, the posterior mean of μ i is used as the estimated expression level of gene i, while the sample mean, ., is referred to as the observed expression level.
As prior knowledge, we assume that if several genes belong to the same operon, in accordance with the RegulonDB annotation, then their expression levels are from a normal distribution, with the mean λ p and the variance τ2. Specifically,
where O p denotes operon p. λ p represents the expression level of the operon p, which is the average of the mean expression levels of all genes within the operon p. τ2 is the within operon variation, and is assumed to be the same across all operons. A non-informative prior is assigned to λ p , that is Pr(λ p ) α 1, to reflect the lack of prior information, and τ2 have vague priors, which are inverse Gamma distributions with the shape and rate parameters equal to 0.01 and 0.01 respectively . If gene i is not in any operon, then
Equation (3) shows that, when borrowing information from the other genes within the same operon, the estimated expression level of the gene i becomes the weighted average of the observed expression level of gene i and the expression level of operon p, given that gene i belongs to operon p. The weights are inversely proportional to the variances. In this model, a key concept is to shrink the observed expression level , towards λ p , the expression level of an operon, based on the knowledge of the operon structure. The degree of shrinkage is determined by the variability of . and λ p . Without incorporating operon information, the estimated expression level would be close to the observed expression level, In the hierarchical model, λ p represents the expression level of operon p, and
where, m p is the number of genes in operon p, and i ∈ O p denotes that gene t is in operon p. Although the posterior distribution was not available in a closed form, we could derive a closed form of the full conditional distribution, and used Markov chain Monte Carlo (MCMC) to simulate the parameters from the posterior distribution. With this closed form expression, the model could be easily coded in R [see Additional file 1] for MCMC simulation using Gibbs sampling . The expression level of gene i is estimated by the posterior mean of μ i , and the genes are ranked by the absolute values of the posterior means of the μ i ' s. Genes with high rankings were designated as differentially expressed (DE) genes.
SAM t statistic
To evaluate the performance of the proposed method in estimating gene expression levels and identifying DE genes, we compared the proposed method to the sample mean and SAM t statistics [1, 3]. Because of its good performance, the SAM t statistic [1, 3] is widely used to rank genes and detect DE genes . We denote the SAM t statistic for the gene i as Z i , then
We conducted three simulation studies to assess the usefulness of our method. The operon structure of the E. coli genome from the RegulonDB database  was used in simulation studies. We randomly chose 100 operons (involving about 340 genes) and assumed that genes from those operons were differentially expressed DE genes. Then we randomly picked a subset of non-operon genes to be DE genes and adjusted the total number of DE genes to 400. Let μ i be the expression level of gene i, for i = 1, 2,..., 4821. For DE genes, μ i ' s were simulated from an equal mixture of N(1, 0.252) and N(– 1,0.252) distributions, and genes within the same operon were from the same component of the mixture distribution. For EE genes, μ i ~ N(0, 0.252). We simulated 4 replicates from a normal distribution for each gene, Y ij ~ N(μ i , ), where Yi,jwas the log ratio of transcript abundances for the gene i on the array j. To provide increasing noise levels for simulations 1, 2 and 3, the σ i 's were simulated from the uniform(0.25, 0.75), uniform( 0.5,1.0), and uniform( 0.75,1.25), respectively.
Estimation of false positives and false negatives
Using the posterior distributions, we can evaluate the FDR for specific number of DE genes. Using Pr(|μ i | > δ'|Y i,j ) to estimate the probability of gene i to be a DE gene, we can estimate the number of false positives for a cut off value k :
Here, the genes are ranked based on the estimated mean expression level μ i . In this study, we set δ = 1,
which corresponding to the commonly used 2-fold cutoff.
The false discovery rate (FDR) for the cut off k can be derived as:
Similarly, the number of false negatives for the cut off k can be calculated as:
Algorithm for Gibbs sampler
The algorithm is implemented below:
Set initial values:
FOR t FROM 1 TO T, draws random samples:
where V i is the sample variance of gene i, and V0 is the median of V i 's. n p is the number of genes in operon p. T is the total number of iteration. To diminish the effect of the initial values, we discard the results from the early iterations (t ≤ T B , where T B is the burn in time). The posterior mean of μ i , of gene i is calculated by:
The authors are grateful to the reviewers for helpful comments. GX is supported by a Merck fellowship, and BMMV was supported in part by a Ford postdoctoral fellowship. This work was supported in part by NIH grant GM066098 (ABK) and HL65462 (WP), and a University of Minnesota AHC faculty research development grant (WP and ABK).
- Tusher VG, Tibshirani R, Chu G: Significance analysis of microarrays applied to the ionizing radiation response. PNAS. 2001, 98 (9): 5116-5121. 10.1073/pnas.091062498. [http://www.pnas.Org/cgi/content/abstract/98/9/5116]PubMedPubMed CentralView ArticleGoogle Scholar
- Baldi P, Long AD: A Bayesian framework for the analysis of microarray expression data: regularized t -test and statistical inferences of gene changes. Bioinformatics. 2001, 17 (6): 509-519. 10.1093/bioinformatics/17.6.509. [http://bioinformatics.oxfordjournals.Org/cgi/content/abstract/17/6/509]PubMedView ArticleGoogle Scholar
- Efron B, Tishirani R, Storey J, Tusher V: Empirical Bayes analysis of a microarray experiment. J Amer Statist Assoc. 2001, 96: 1151-1160. 10.1198/016214501753382129.View ArticleGoogle Scholar
- Pan W: A comparative review of statistical methods for discovering dierentially expressed genesin replicated microarray experiments. Bioinformatics. 2002, 18 (4): 546-554. 10.1093/bioinformatics/18.4.546. [http://bioinformatics.oxfordjournals.Org/cgi/content/abstract/18/4/546]PubMedView ArticleGoogle Scholar
- Broet P, Richardson S, Radvanyi F: Bayesian Hierarchical Model for Identifying Changes in Gene Expression from Microarray Experiments. Journal of Computational Biology. 2002, 9 (4): 671-683. 10.1089/106652702760277381. [http://www.liebertonline.com/doi/abs/10.1089/106652702760277381]PubMedView ArticleGoogle Scholar
- Kendziorski CM, Newton MA, Lan H, Gould MN: On parametric empirical Bayes methods for comparing multiple groups using replicated gene expression profiles. Statistics in Medicine. 2003, 22 (24): 3899-3914. 10.1002/sim.1548.PubMedView ArticleGoogle Scholar
- Newton MA, Noueiry A, Sarkar D, Ahlquist P: Detecting differential gene expression with a semiparametric hierarchical mixture method. Biostat. 2004, 5 (2): 155-176. 10.1093/biostatistics/5.2.155. [http://biostatistics.oxfordjournals.Org/cgi/content/abstract/5/2/155]View ArticleGoogle Scholar
- Lonnstedt I, Speed T: Replicated microarray data. Statist Sinica. 2002, 12: 31-46.Google Scholar
- Lewin A, Richardson S, Marshall C, A G, Aitman T: Bayesian Modelling of Differential Gene Expression. Biometrics. 2005,http://www.bgx.org.uk/papers.html, ,Google Scholar
- Liu D, Parmigiani G, Caffo B: Screening for Differentially Expressed Genes: Are Multilevel Models Helpful?. Johns Hopkins University, Dept. of Biostatistics Working Papers. 2004, [http://www.bepress.com/jhubiostat/paper34]Google Scholar
- Miller JH, Reznikoff WS: The operon. 1978, Cold Spring Harbor, NY: Cold Spring Harbor Laboratory PressGoogle Scholar
- Khodursky AB, Peter BJ, Cozzarelli NR, Botstein D, Brown PO, Yanofsky C: DNA microarray analysis of gene expression in response to physiological and genetic changes that affect tryptophan metabolism in Escherichia coli. PNAS. 2000, 97 (22): 12170-12175. 10.1073/pnas.220414297. [http://www.pnas.org/cgi/content/abstract/97/22/12170]PubMedPubMed CentralView ArticleGoogle Scholar
- Courcelle J, Khodursky A, Peter B, Brown PO, Hanawalt PC: Comparative Gene Expression Profiles Following UV Exposure in Wild-Type and SOS-Deficient Escherichia coli. Genetics. 2001, 158: 41-64. [http://www.genetics.Org/cgi/content/full/158/l/41]PubMedPubMed CentralGoogle Scholar
- Moreno-Hagelsieb G, Trevino V, Perez-Rueda E, Smith TF, Collado-Vides J: Transcription unit conservation in the three domains of life: a perspective from Escherichia coli. Trends Genet. 2001, 17 (4): 175-7. 10.1016/S0168-9525(01)02241-7.PubMedView ArticleGoogle Scholar
- Salgado H, Moreno-Hagelsieb G, Smith TF, Collado-Vides J: Operons in Escherichia coli: Genomic analyses and predictions. PNAS. 2000, 97 (12): 6652-6657. 10.1073/pnas.110147297. [http://www.pnas.org/cgi/content/abstract/97/12/6652]PubMedPubMed CentralView ArticleGoogle Scholar
- Moreno-Hagelsieb G, Collado-Vides J: A powerful non-homology method for the prediction of operons in prokaryotes. Bioinformatics. 2002, S329-36. 18 Suppl 1(NIL)
- Ermolaeva MD, White O, Salzberg SL: Prediction of operons in microbial genomes. Nucleic Acids Res. 2001, 29 (5): 1216-21. 10.1093/nar/29.5.1216.PubMedPubMed CentralView ArticleGoogle Scholar
- Jacob E, Sasikumar R, Nair KNR: A fuzzy guided genetic algorithm for operon prediction. Bioinformatics. 2005, 21 (8): 1403-7. 10.1093/bioinformatics/bti156.PubMedView ArticleGoogle Scholar
- Westover BP, Buhler JD, Sonnenburg JL, Gordon JI: Operon prediction without a training set. Bioinformatics. 2005, 21 (7): 880-8. 10.1093/bioinformatics/bti123.PubMedView ArticleGoogle Scholar
- Jeong KS, Ahn J, Khodursky AB: Spatial patterns of transcriptional activity in the chromosome of Escherichia coli. Genome Biology. 2004, 5: R86-10.1186/gb-2004-5-11-r86.PubMedPubMed CentralView ArticleGoogle Scholar
- Sabatti C, Rohlin L, Oh MK, Liao JC: Co-expression pattern from DNA microarray experiments as a tool for operon prediction. Nucleic Acids Res. 2002, 30 (13): 2886-93. 10.1093/nar/gkf388.PubMedPubMed CentralView ArticleGoogle Scholar
- Bockhorst J, Craven M, Page D, Shavlik J, Glasner J: A Bayesian network approach to operon prediction. Bioinformatics. 2003, 19 (10): 1227-35. 10.1093/bioinformatics/btg147.PubMedView ArticleGoogle Scholar
- Wren JD, Yao M, Langer M, Conway T: Simulated annealing of microarray data reduces noise and enables cross-experimental comparisons. DNA Cell Biol. 2004, 23 (10): 695-700. 10.1089/dna.2004.23.695.PubMedView ArticleGoogle Scholar
- Sangurdekar DP, Srienc F, Khodursky AB: A classification based framework for quantitative description of large-scale microarray data. Genome Biology. 2006, 7 (4): R32-10.1186/gb-2006-7-4-r32.PubMedPubMed CentralView ArticleGoogle Scholar
- Macnab RM: Genetics and biogenesis of bacterial flagella. Annu Rev Genet. 1992, 131-58. 10.1146/annurev.ge.26.120192.001023. 26(NIL)
- Benjamini Y, Hochberg Y: Controlling the false discovery rate: A practical and powerful approach to multiple testing. J R Stat Soc B. 1995, 57: 289-300.Google Scholar
- Storey JD, Tibshirani R: Statistical significance for genomewide studies. PNAS. 2003, 100 (16): 9440-9445. 10.1073/pnas.1530509100. [http://www.pnas.org/cgi/content/abstract/100/16/9440]PubMedPubMed CentralView ArticleGoogle Scholar
- Lercher MJ, Blumenthal T, Hurst LD: Coexpression of neighboring genes in Caenorhabditis elegans is mostly due to operons and duplicate genes. Genome Res. 2003, 13 (2): 238-43. 10.1101/gr.553803.PubMedPubMed CentralView ArticleGoogle Scholar
- Blumenthal T, Gleason KS: Caenorhabditis elegans operons: form and function. Nat Rev Genet. 2003, 4 (2): 112-20. 10.1038/nrg995.PubMedView ArticleGoogle Scholar
- Blumenthal T: Operons in eukaryotes. Brief Funct Genomic Proteomic. 2004, 3 (3): 199-211. 10.1093/bfgp/3.3.199.PubMedView ArticleGoogle Scholar
- Cohen BA, Mitra RD, Hughes JD, Church GM: A computational analysis of whole-genome expression data reveals chromosomal domains of gene expression. Nat Genet. 2000, 26 (2): 183-6. 10.1038/79896.PubMedView ArticleGoogle Scholar
- Kruglyak S, Tang H: Regulation of adjacent yeast genes. Trends Genet. 2000, 16 (3): 109-11. 10.1016/S0168-9525(99)01941-1.PubMedView ArticleGoogle Scholar
- Roy PJ, Stuart JM, Lund J, Kim SK: Chromosomal clustering of muscle-expressed genes in Caenorhabditis elegans. Nature. 2002, 418 (6901): 975-9.PubMedGoogle Scholar
- Boutanaev AM, Kalmykova AI, Shevelyov YY, Nurminsky DI: Large clusters of co-expressed genes in the Drosophila genome. Nature . 2002, 420 (6916): 666-9. 10.1038/nature01216.PubMedView ArticleGoogle Scholar
- Spellman PT, Rubin GM: Evidence for large domains of similarly expressed genes in the Drosophila genome. J Biol. 2002, 1: 5-10.1186/1475-4924-1-5.PubMedPubMed CentralView ArticleGoogle Scholar
- Li Q, Lee BTK, Zhang L: Genome-scale analysis of positional clustering of mouse testis-specific genes. BMC Genomics. 2005, 6: 7-10.1186/1471-2164-6-7.PubMedPubMed CentralView ArticleGoogle Scholar
- Caron H, van Schaik B, van der Mee M, Baas F, Riggins G, van Sluis P, Hermus MC, van Asperen R, Boon K, Voute PA, Heisterkamp S, van Kampen A, Versteeg R: The human transcriptome map: clustering of highly expressed genes in chromosomal domains. Science. 2001, 291 (5507): 1289-92. 10.1126/science.1056794.PubMedView ArticleGoogle Scholar
- Versteeg R, van Schaik BDC, van Batenburg MF, Roos M, Monajemi R, Caron H, Bussemaker HJ, van Kampen AHC: The human transcriptome map reveals extremes in gene density, intron length, GC content, and repeat pattern for domains of highly and weakly expressed genes. Genome Res. 2003, 13 (9): 1998-2004. 10.1101/gr.1649303.PubMedPubMed CentralView ArticleGoogle Scholar
- Yager TD, Dempsey AA, Tang H, Stamatiou D, Chao S, Marshall KW, Liew CC: First comprehensive mapping of cartilage transcripts to the human genome. Genomics. 2004, 84 (3): 524-35. 10.1016/j.ygeno.2004.05.006.PubMedView ArticleGoogle Scholar
- Salgado H, Gama-Castro S, Martinez-Antonio A, Diaz-Peredo E, Sanchez-Solano F, Peralta-Gil M, Garcia-Alonso D, Jimenez-Jacinto V, Santos-Zavaleta A, Bonavides-Martinez C, Collado-Vides J: RegulonDB (version 4.0): transcriptional regulation, operon organization and growth conditions in Escherichia coli K-12. Nucl Acids Res. 2004, 32 (90001): D303-306. 10.1093/nar/gkh140. [http://nar.oxfordjournals.org/cgi/content/full/32/suppLl/D303]PubMedPubMed CentralView ArticleGoogle Scholar
- NCBI Gene Expression Omnibus. [http://www.ncbi.nlm.nih.gov/geo/]
- Yang YH, Dudoit S, Luu P, Lin DM, Peng V, Ngai J, Speed TP: Normalization for cDNA microarray data: a robust composite method addressing single and multiple slide systematic variation. Nucl Acids Res. 2002, 30 (4): e15-10.1093/nar/30.4.e15. [http://nar.oxfordjournals.Org/cgi/content/full/30/4/el5]PubMedPubMed CentralView ArticleGoogle Scholar
- Carlin B, Louis T: Bayes and Empirical Bayes Methods for Data Analysis. 2000, Boca Raton, FL: Chapman and Hall/CRC Press 2000View ArticleGoogle Scholar
- Gelfand A, Smith A: Sampling Based Approaches to Calculating Marginal Densities. Journal Amer Stat Assoc. 1990, 85: 398-409. 10.2307/2289776.View ArticleGoogle Scholar
- Xie Y, Jeong KS, Pan W, Khodursky A, Carlin BP: A case study on choosing normalization methods and test statistics for two-channel microarray data. Comp Fund Genom. 2004, 5: 432-444. 10.1002/cfg.416.View ArticleGoogle Scholar
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