The effects of normalization on the correlation structure of microarray data
 Xing Qiu^{1},
 Andrew I Brooks^{2},
 Lev Klebanov^{1, 3} and
 Andrei Yakovlev^{1}Email author
https://doi.org/10.1186/147121056120
© Qiu et al; licensee BioMed Central Ltd. 2005
Received: 16 December 2004
Accepted: 16 May 2005
Published: 16 May 2005
Abstract
Background
Stochastic dependence between gene expression levels in microarray data is of critical importance for the methods of statistical inference that resort to pooling teststatistics across genes. It is frequently assumed that dependence between genes (or tests) is suffciently weak to justify the proposed methods of testing for differentially expressed genes. A potential impact of betweengene correlations on the performance of such methods has yet to be explored.
Results
The paper presents a systematic study of correlation between the tstatistics associated with different genes. We report the effects of four different normalization methods using a large set of microarray data on childhood leukemia in addition to several sets of simulated data. Our findings help decipher the correlation structure of microarray data before and after the application of normalization procedures.
Conclusion
A longrange correlation in microarray data manifests itself in thousands of genes that are heavily correlated with a given gene in terms of the associated tstatistics. By using normalization methods it is possible to significantly reduce correlation between the tstatistics computed for different genes. Normalization procedures affect both the true correlation, stemming from gene interactions, and the spurious correlation induced by random noise. When analyzing real world biological data sets, normalization procedures are unable to completely remove correlation between the test statistics. The longrange correlation structure also persists in normalized data.
Keywords
Background
There are two major methodological problems that deal with the issue of stochastic dependence between gene expression signals in microarray data. The first arises naturally when adjustments for multiplicity of tests are made by pooling across genes (or tests) in an effort to find differentially expressed genes in twosample comparisons. The empirical Bayes methodology in the nonparametric [1–3] and parametric formulations [4, 5], and closely related methods exploiting a twocomponent mixture model [6–8] represent typical examples. The common feature of such methods is that a test statistic (measure of differential expression) is first calculated for each gene to account for biological variability and then all the statistics (or the associated pvalues) are pooled together and treated as a sample from which to estimate the sampling distribution of this statistic, the false discovery rate (FDR), qvalues, etc. The same kind of pooling is typically used in maximum likelihood inference from microarray data [9, 10] and some other methods of testing for differential expression of genes.
In all such approaches, the stochastic dependence between gene expression values or test statistics is a nuisance that hinders their application. The independence assumption is frequently invoked when building a theoretical foundation for a particular method of statistical inference. Some authors (e.g., [11]) allow for dependence between differentially expressed genes while assuming stochastic independence of those genes that do not change their expression between the two conditions under study. The biological rationale for such a hypothesis is unclear, because the normally functioning genes are involved in numerous biochemical pathways much like the altered ones.
The stochastic dependence between expression levels and thus between the associated test statistics is really a serious problem. It may cause high variability of statistical estimators and even deteriorate their consistency. To obtain theoretical results it is frequently assumed that weak or almost sure convergence holds for an empirical distribution function constructed from the data pooled across genes (see, i.e. [12, 13]). However, this assumption is diffcult to validate biologically so that the required convergence to the true distribution function is always questionable; it may or may not be the case depending on the type and strength of stochastic dependence.
Storey [12] advocates the assumption of weak dependence between teststatistics when discussing some concerns raised in the paper by Ge, Dudoit, and Speed (hereafter abbreviated by GDS) [14]. It is worth quoting his line of reasoning at length:
"I hypothesize that the most likely form of dependence between the genes encountered in DNA microarrays is weak dependence, and more specifically, "clumpy dependence"; that is, the measurements on the genes are dependent in small groups, each group being independent of the others. There are two reasons that make clumpy dependence likely. The first is that genes tend to work in pathways, that is, small groups of genes interact to produce some overall process. This can involve just a few to 50 or more genes. This would lead to a clumpy dependence in the pathwayspecific noise in the data. The second reason is that there tends to be crosshybridization in DNA microarrays. In other words, the signals between two genes can cross because of molecular similarity at the sequence level. Crosshybridization would only occur in small groups, and each group would be independent of the others."
This hypothesis does not seem plausible from a biological standpoint because of the pleiotropic character of gene function: one gene participates in multiple molecular pathways. However, the possibility that it may approximately be true for all practical purposes cannot be ruled out. There are two key words in the above quotation: "small groups" and "weak dependence". Whether or not such groups are small and stochastic dependence is suffciently weak can be deciphered only from real world data. To the best of our knowledge, no attempt has been made so far to systematically study dependence structures in microarray data using large data sets. In this connection we would like to continue quoting from [12]: "Many assumptions that have been made for modeling microarray data have yet to be verified. Hopefully evidence either for or against these assumptions will emerge... GDS have stressed the dependence between the genes... I leave it as a challenge to them to provide evidence from real microarray data that the aforementioned assumptions do not hold. I have not been able to find it myself". In the present paper, we take the first step in this direction by conducting an empirical study of the correlations between test statistics associated with different genes.
The second research area where the dependence between gene expression levels plays a crucial role is the discovery (reverse engineering) of molecular pathways and networks from microarray data [15]. A popular approach to pathway reconstruction is based on the sample correlation coeffcient or mutual information measures that are deemed to characterize interactions between genes via their products. These measures of interaction are computed from gene expression values observed across various experimental conditions. The snag here is that strong correlations in the raw (not normalized but background corrected) expression data may be induced by an arrayspecific technological noise, thereby producing numerous falsepositive edges in the corresponding graph representing the underlying structure of a given pathway or network. However, if the data are normalized before the analysis, then the correlation structure of expression signals may be partially destroyed by the normalization procedure so that many edges in the resultant graph may be missing. The same applies to clustering techniques that utilize information on pairwise dependencies between the genes. The problem is less pressing where causal inference is possible from gene perturbation experiments. Although the present paper does not have a direct bearing on such settings, our results suggest that associative networks built on microarray data alone may have little to do with biological reality. The problem merits careful investigation in order to make the reverse engineering of this type more credible.
The present paper is focused on the correlations between teststatistics associated with expression signals produced by each gene and the effects of normalization procedures on these correlations. We limit our consideration to the tstatistic which is the most popular choice in microarray data analysis. Normalization is intended to mitigate the effect of technological noise that is inherent in microarray data. Normalization procedures tend to reduce the variability of original microarray data (Park et al, [16]), however no study has been carried out to assess the effect of such procedures on the correlation structure of microarray data in general and the correlation of tstatistics in particular. In a methodological study such as ours, it is a great advantage to have access to a large data set involving hundreds of arrays. We used the St. Jude Children's Research Hospital (SJCRH) Database on childhood leukemia which falls into this category. Computer simulations provide the necessary, albeit not very realistic, control where the actual model is known and arbitrarily large samples can be generated for testing various methodologies.
Results
The design of our study is presented in the Methods section. This design allows us to compute the tstatistics (across arrays) for each gene and each pair of subsamples. This computation results in 15 values (corresponding to the 15 pairs of subsamples) of the tstatistic associated with each gene. Then we compute the sample correlation coeffcients between the tstatistics thus obtained for every pair of genes. The resulting coeffcients are summarized in the form of a histogram. We interpret such histograms as pertinent summary characteristics and not as estimators of some population distribution densities. We also look at preselected individual genes to determine the range of their correlation with all other genes. This range can be characterized by the number of gene pairs formed by a given gene with the correlation coeffcient exceeding some threshold level. We adopt the value of 0.5 as such a threshold.
Using these tools we attempt to answer the following questions:

What is the (pairwise) correlation structure of the tstatistic in a large population of genes?

What is the impact of normalization procedures on this structure?

What is the impact of normalization procedures on the number of highly correlated pairs formed by a given gene?
The effects of three normalization procedures (GEO, RANK, and QUANT, as defined in the Methods section) are shown in Figures 1B–1D. Figure 1E presents an ideal case where the tstatistics were obtained from independent normally distributed data (see the Methods section for explanations) produced by simulations (SIMU1). In this case, the proportion of gene pairs with correlation coeffcients greater than 0.5 is only 1.5%. While the normalization procedure GEO destroys a large proportion of correlation, the procedures RANK and QUANT outperform it as far as the reduction of betweengene dependence is concerned. The effects of the latter two procedures are largely similar. The procedure RANK reduces the proportion of correlation coeffcients greater than 0.5 to 4.3%, while the procedure QUANT reduces this proportion to 7.2%. For comparison, this indicator is equal to 14% for GEO. Thus the procedure RANK has the strongest effect on the correlation structure. Figure 1 in the Additional Material Files [see the file "Additional File 1"] shows essentially the same effect for randomly selected nonoverlapping pairs of genes.
The effect of the quantile normalization for the SIMU3N, shown in Figure 3 in the Additional Material Files [see "Additional File 2"], deserves special discussion. Recall that each gene in the data set SIMU3N correlates only with a distinct group of genes termed a clump. Even if the genes involved in the same clump are heavily correlated, the average (over all pairs of genes) correlation coeffcient may still be quite low. When a uniformly distributed multiplicative random noise is imposed on each array, the genes pertaining to different clumps become highly correlated. The noise strengthens the intraclump correlation as well. Recall that the clumpy structure of simulated data serves as a simplistic model of gene interactions within distinct pathways. As seen in Figure 3 [see "Additional File 2"], the normalization procedure QUANT is not nearly as effective as in the case of the SIMU2N data. This procedure does not eradicate the overall correlation between genes in the SIMU3N data. In this sense, the effects of normalization seen in the SIMU3N and in real biological data look similar.
Longrange correlation analysis for the SIMU2N data.
Gene Label  GEO  QUANT  RANK  SIMU2N 

1  743  746  741  12558 
2  754  750  756  12558 
3  723  723  721  12558 
4  705  698  718  12558 
5  736  734  754  12558 
6  751  763  765  12558 
7  702  695  709  12558 
8  667  665  679  12558 
9  747  747  759  12558 
10  728  730  736  12558 
11  713  717  713  12558 
12  696  699  685  12558 
13  743  750  762  12558 
14  725  721  733  12558 
15  691  691  740  12558 
16  789  789  799  12558 
17  724  725  669  12558 
18  716  712  722  12558 
19  762  762  720  12558 
20  676  673  708  12558 
Mean  724.6  724.5  729.5  12558 
STD  30.1  31.8  31.9  0 
Longrange correlation analysis for the SIMU3N data.
Gene Label  GEO  QUANT  RANK  SIMU3N 

1  483  520  512  12297 
2  471  582  591  10656 
3  436  523  614  12506 
4  644  643  744  11031 
5  677  739  765  11320 
6  610  543  570  12413 
7  612  863  788  12429 
8  802  727  711  12077 
9  1743  1406  1077  11898 
10  975  895  920  12001 
11  1352  1330  1543  12453 
12  670  707  686  12480 
13  1874  1849  1890  6913 
14  1858  1765  1808  9371 
15  1925  1790  1974  12469 
16  1792  1718  1796  12520 
17  1764  1526  1679  12499 
18  1769  1684  1821  12509 
19  1476  1300  1569  12514 
20  2223  2307  2148  12507 
Mean  1207.8  1170.9  1210.3  11743.2 
STD  617.3  557.5  576.5  1402 
Longrange correlation analysis for the SJCRH data.
Gene Label  GEO  QUANT  RANK  raw data 

1  5644  462  494  12481 
2  7330  3175  1431  12486 
3  4189  1480  2062  12496 
4  5218  2728  1548  12493 
5  8169  1888  1064  12451 
6  8140  956  1162  12482 
7  323  1169  839  12480 
8  6774  1479  839  12497 
9  7676  1832  2140  12390 
10  8234  794  1440  12384 
11  7652  930  466  12498 
12  8266  1329  708  12476 
13  8197  1343  2045  12391 
14  7422  2118  2513  12501 
15  1588  1467  1011  12494 
16  7861  1931  1133  12429 
17  1292  1477  1445  12489 
18  6389  2949  1456  12481 
19  7359  490  514  12469 
20  4384  970  787  12488 
Mean  6105.4  1548.4  1254.9  12467.8 
STD  2545 2512  756  589.5  38.2 
Consider first the results obtained with simulated data. Each of the twenty initiator genes selected from SIMU2N form exactly 12,558 highly correlated pairs. When applied to the SIMU2N data, the normalization procedures RANK and QUANT bring this number down to 700 on average (see Table 1). The variability in the size of this set of genes is low. For example, the number of highly correlated genes ranges from 661 to 794 after the application of the QUANT procedure. Both procedures indiscriminately reduce the true (intrinsic) correlation and its spurious (nuisance) counterpart. Although less effective, the procedure GEO does a similar job.
The results for the SIMU3N data are different (see Table 2). While the number of highly correlated gene pairs tends to decrease significantly for each of the twenty initiator genes, the size of this effect depends on the group of genes from which the initiator gene was chosen. This increases the variability of the number of highly correlated pairs remaining after normalization. For the QUANT method the range is from 526 to 2,368 showing that the remaining correlation extends far beyond the specified clumpy structure.
We then selected 20 initiator genes in the SJCRH data set representing real biological data. The number of highly correlated pairs formed by these genes before normalization ranges from 12,384 to 12,501, which is a very narrow range indeed. As is seen in Table 3, the procedure GEO does not destroy the correlation effectively; it leaves huge numbers (up to 8,266) of highly correlated gene pairs. The rank normalization results in much smaller numbers of highly correlated genes that range from the lowest of 494 to the highest of 2,513. The average is 1,255, which is about twice as much as we get from any normalized SIMU2N data. The variability is also very high, resembling a clumpy effect seen in the SIMU3N set. We do not consider this similarity as evidence for a clumpy structure of microarray data, but the results in Table 3 suggest that, if such a structure exists, an average clump should be expected to involve at least an order of magnitude more genes than the clump size postulated by Storey [12].
Another interesting finding in Table 3 is that the quantile normalization tends to leave more highly correlated genes in comparison to the rank normalization. This is contrary to our expectations based on the comparisons of correlation histograms reported above. The effect of the QUANT is also more variable than that of the RANK, which is another dissimilarity of practical importance. Leaving aside the fact that the RANK procedure is applied to gene expressions, while the QUANT works at the probe feature level, the difference between the two normalization methods is that we replace entries in an array by their ranks in the former case and by in the latter. Recall that is the average of entries having the same rank over all arrays. Obviously, the QUANT preserves more quantitative information in the data than does the RANK procedure. This explains why the result of the rank normalization is less variable.
The effect of the normalization QUANT on the distribution of the tstatistics across the genes for the actual and simulated data is shown in Figures 1, 2, 3, 4 included in the Additional Material Files [see the file "Additional File 3"]. From Figures 2 and 3, it is clear that, when applied to the simulated data SIMU2 and SIMU2N, this procedure makes the distribution of tstatistics similar to that in the ideal case shown in Figure 1 [see "Additional File 3"]. However, the effect appears to be somewhat less satisfactory with real data, especially in the tail regions of the resultant distribution of tstatistics.
The results shown in this section are obtained with a single initial random split of the pooled set of arrays into two groups. We have conducted several such splits in this study. All the abovedescribed effects are highly reproducible, and reporting the results for other splits in the paper is not warranted.
Discussion
It follows from our observations that normalization procedures are capable of destroying a significant part of correlations between gene expression signals and associated teststatistics. In doing so, they affect both the spurious correlation induced by the noise and the true correlation that reflects gene interactions. The clumpy structure (involving relatively large clumps of genes) of the SIMU3N data set is more resistant to this effect than the SIMU2N data. This is even more so for real biological data. The weaker effect of normalization seen in the SJCRH data indicates that the actual noise structure may be more complicated than assumed in the simulation studies (multiplicative arrayspecific random effect model). A clumpy structure of gene expression signals may also play a role in this phenomenon. This observation explains why it is so diffcult to remove correlations from the data.
The destructive effect of normalization procedures on pairwise correlations in microarray data is good news for the methods of statistical inference that resort to "pooling across genes". However, it remains unclear whether or not the remaining correlation may still be substantial enough to invalidate such methods by affecting important properties of statistical estimators and tests. The problem invites further investigation. However, we would like to present an experiment specially designed to address the consistency question mentioned in the Background section.
 1.
Select randomly 100 genes and compute the arithmetic (sample) mean of the tstatistics across these genes for each pair of subsamples.
 2.
Compute the standard deviation of the sample mean across the 15 pairs of subsamples.
 3.
Select randomly 100 from the remaining genes and compute the arithmetic mean for the 200 genes for each pair of subsamples.
 4.
Compute the standard deviation from the sample means resulted from the previous step.
 5.
Continue until the set of all genes is exhausted.
 6.
Plot the estimated standard deviation of the sample mean as a function of the number of genes involved in each step of the algorithm.
 7.
Repeat the procedure k times to generate k trajectories of the standard deviation of the sample mean.
The results of one such experiment are given in Figure 3. It is known that the sample mean is an unbiased and consistent estimator for the true mean value in the case of independent and identically distributed observations. This case is represented by Curve 3 generated by simulations. It is clear that the standard deviation decreases very rapidly and tends to zero with increasing the number of genes. However, the same is not true for the biological data. For the raw data, the standard deviation does not show a distinct tendency to decrease (Curve 1). When the data are normalized using the quantile normalization procedure, the standard deviation first drops and then stabilizes at an approximately constant level, no matter how many (up to 12,500) genes are involved in its estimation (see Curve 2). This is clearly the effect of (longrange) correlation between the tstatistics associated with different genes. The pattern seen in Figure 3 was highly reproducible across k = 20 experiments with different random starts. If the standard deviation of an unbiased estimator tends to zero, this estimator is consistent. This is the case for Curve 3 but not quite so for Curve 2. While not a rigorous disproof of consistency of the sample mean in this case, the pattern seen in Curve 2 suggests that the estimator is likely to converge to a random variable (with the variance greater than zero) rather than to the true parameter to be estimated. This is definitely not a good sign for estimation procedures based on pooling across genes such as those built in the empirical Bayes methodology.
The observed effect of normalization procedures is definitely bad news for the associative network reconstruction from gene expression data. Unless further technological advancements result in a significant reduction of the noise in microarray data, this kind of analysis will continue producing unreliable inferences. To normalize, or not to normalize: that is the question to which no scientifically sound answer is currently known as far as this kind of reverse engineering is concerned. Although limited to cell cultures, the causal inference from gene perturbation (disruption and overexpression) experiments seems to be the only solid alternative. From this standpoint the observations reported in the present paper add to the concerns expressed by several investigators regarding how much confidence to place in the thousands of papers already published using microarray technology [17].
Conclusion
The present paper provides quantitative insight into correlation between the tstatistics associated with different genes. This study leads us to conclude that:

There is a longrange correlation in microarray data manifesting itself in a huge number of genes that are heavily correlated with a given gene in terms of the associated tstatistics.

Using normalization of microarray data it is possible to significantly reduce correlation between the tstatistics computed for different genes.

Normalization procedures affect both the true correlation, stemming from gene interactions, and the spurious correlation induced by random noise.

It is likely that some noise effects represent nonmonotone transformations of the underlying gene expression signals because even the rank normalization does not make the tstatistics independent when applied to the biological data.

Even the most effcient normalization procedures are unable to completely remove correlation between the tstatistics associated with different genes in biological data. Furthermore, the longrange correlation structure persists in normalized data. This remaining correlation may be strong enough to deteriorate consistency of statistical estimators built from measurements on the genes.
Methods
Study design and biological data
There are 335 arrays (Affymetrix, Santa Clara, CA) in the SJCRH data set, each array representing N = 12, 558 genes. Each gene is represented in the data set by the logarithm of its expression level. The data are publicly available on the following website: [18]. The SJCRH data include the information on gene expression in normal blood and various types of childhood leukemia. The raw (background corrected but not normalized) expression data were generated by the output of the Bioconductor RMA (Robust MultiArray Average) procedure when choosing the option: normalization = false. Since our focus was on purely methodological problems, we pooled all the available arrays together and shuffed the pooled sample. After randomly choosing and dropping 5 arrays (to make all subsamples of the same size), the pooled sample was randomly split into 30 parts, each containing 11 arrays.
Then 15 pairs of the array samples were arranged and the corresponding 15 tstatistics were computed for each gene, thereby mimicking 15 twosample comparisons under the null hypothesis of no differential expression. As a result, each gene was associated with 15 values of the tstatistic so that the Pearson correlation coeffcient between the tstatistics thus derived could be computed for any pair of genes. The output of the abovedescribed series of procedures is a 12, 558 × 15 matrix of tstatistics and the associated vector of correlation coeffcients for all pairs of genes. We proceeded through the same sequence of operations when analyzing normalized and simulated data sets.
Simulated data
 1.
SIMU1: Every element x_{ ij }, i = 1, ..., 12, 558; j = 1, ..., 335 in SIMU1 represents logintensity of expression of the i th gene from the j th array. The independent and identically distributed random variables x_{ ij }are generated from the standard normal distribution. This implies that the original expression signals are modeled as lognormally distributed random variables but we used their logarithms in our computations. This data set was used to illustrate the correlation analysis under independence of gene expression levels.
 2.
SIMU2 is a 12, 558 × 335 random matrix that models an exchangeable correlation structure. The entries in this matrix are normal random variables with mean zero and unit variance. The entries from different columns are independent, while the correlation coeffcient between any two elements x_{ ij }of the same column is equal to 0.8.
 3.
SIMU2N is a data set based on SIMU2. First we generate a 335dimensional random vector A. The elements of A are independent and identically distributed. The marginal distri bution of every element a_{ ij }of A is uniform over the interval [5, 10]. This random vector is used to model an arrayspecific noise. We then define y_{ ij }to be x_{ ij }+ a_{ j }, where x_{ ij }is the ijth entry in the SIMU2 data set. The new matrix Y = {y_{ ij }} represents the data SIMU2N
 4.
SIMU3 is a 12,550 × 335 matrix. The 12,550 rows (genes) are divided into ten groups of genes, each containing 1,255 rows. If two genes are both from the kth group (gene numbers 100·(k1)+1 through 100·k), for k = 1, 2, ..., 10, then the correlation coeffcient between them equals . Any two genes pertaining to different groups are stochastically independent.
 5.
SIMU3N is the same as the SIMU3 data set but with an added noise. An arrayspecific multiplicative and uniformly distributed noise is modeled exactly as in the SIMU2N data.
Normalization methods
Suppose there are M arrays of length N, and we represent the corresponding logintensities as an N × M matrix X such that each array is represented by a column in X. In this work, we used the following normalization methods:
1. Geometric mean normalization GEO
If the arrayspecific random noise is multiplicative then a reasonable way to remove it from the expression values is to divide each element of the data matrix by the geometric mean over all gene expression signals on the array to which this element belongs. Szabo et al [20] discuss conditions under which this method is a valid one for testing twosample hypotheses with microarray data.
2. Rank normalization RANK
This method was proposed by Tsodikov et al [19] and discussed further in [20]. In accordance with their suggestion, we first obtain a vector X^{ sort }by arranging all gene expression signals for the same array in increasing order. Next we replace every entry in this array by its position (rank) in X^{ sort }counted from the smallest value. The idea behind this method is that ranks are invariant to any monotone transformation, implying a much more general model for the technological noise than the multiplicative arrayspecific random effect model.
3. Quantile normalization QUANT
As discussed in [21, 22], this method is motivated by the idea that a quantilequantile plot shows that the distribution of M data vectors is the same if the plot is a straight line in the direction of unit vector but it is not the same otherwise. So we could make a set of data to have the same distribution if we projected the points of an Mdimensional quantile plot onto the diagonal. Much like as with the RANK method, this approach is applied to genes rather than arrays. We refer the reader to [21, 23] for more details. When working with the SJCRH data, this method was applied to probe feature level measurements. When working with simulated data, the method was applied to the levels of gene expression directly by processing them in exactly the same way.
Declarations
Acknowledgements
We are grateful to anonymous reviewers whose comments have helped us improve the manuscript. We thank our colleague Cristine Brower for technical assistance. The research is supported in part by NIH Grant GM075299 and Czech Ministry of Education Grant MSM 113200008.
Authors’ Affiliations
References
 Efron B, Tibshirani R, Storey JD, Tusher V: Empirical Bayes analysis of a microarray experiment. J Amer Statist Assoc 2001, 96: 1151–1160. 10.1198/016214501753382129View ArticleGoogle Scholar
 Efron B: Robbins, empirical Bayes and microarrays. Ann Statist 2003, 31: 366–378. 10.1214/aos/1051027871View ArticleGoogle Scholar
 Efron B: Largescale simultaneous hypothesis testing: The choice of a null hypothesis. J Amer Statist Assoc 2004, 99: 96–104.View ArticleGoogle Scholar
 Newton MA, Kendziorski CM, Richmond CS, Blattner FR, Tsui KW: On differential variability of expression ratios: Improving statistical inference about gene expression changes from microarray data. J of Comput Biol 2000, 8(1):37–52. 10.1089/106652701300099074View ArticleGoogle Scholar
 Newton MA, Kendziorski CM: Parametric empirical Bayes methods for microarrays. In The Analysis of Gene Expression Data. Edited by: Parmigiani G, Garrett ES, Irizarry RA, Zeger SL. Springer, New York; 2003:254–271.View ArticleGoogle Scholar
 McLachlan GJ, Do KA, Ambroise C: Analyzing Microarray Gene Expression Data. Wiley, New York; 2004.View ArticleGoogle Scholar
 Dalmasso C, Broet P, Moreau T: A simple procedure for estimating the false discovery rate. Bioinformatics 2004.Google Scholar
 Broet P, Lewin A, Richardson S, Dalmasso C, Magdalenat H: A mixture modelbased strategy for selecting genes in multiclass response microarray experiments. Bioinformatics 2004, 2562–2571. 10.1093/bioinformatics/bth285Google Scholar
 Ideker T, Thorsson V, Seigel AF, Hood LE: Testing for differentially expressed genes by maximum likelihood analysis of microarray data. J Comput Biol 2000, 7(6):805–817. 10.1089/10665270050514945View ArticlePubMedGoogle Scholar
 Segal E, Wang H, Koller D: Discovering molecular pathways from protein interactions and gene expression data. Bioinformatics 2003, 19: i264i272. 10.1093/bioinformatics/btg1037View ArticlePubMedGoogle Scholar
 Tsai CA, Hsueh HM, Chen JJ: Estimation of false discovery rates in multiple testing: application to gene microarray data. Biometrics 2003, 59: 1071–1081. 10.1111/j.0006341X.2003.00123.xView ArticlePubMedGoogle Scholar
 Storey JD: Comment on 'Resamplingbased multiple testing for DNA microarray data analysis' by Ge, Dudoit, and Speed. Test 2003, 12: 1–77.View ArticleGoogle Scholar
 Storey JD, Taylor JE, Siegmund D: Strong control, conservative point estimation and simultaneous conservative consistency of false discovery rates: a unified approach. J R Statist Soc B 2004, 66: 187–205. 10.1111/j.14679868.2004.00439.xView ArticleGoogle Scholar
 Ge Y, Dudoit S, Speed TP: Resamplingbased multiple testing for DNA microarray data analysis. TEST 2003, 12(1):1–44.View ArticleGoogle Scholar
 D'haeseller P, Liang S, Somogyi R: Genetic network inference: from coexpression clustering to reverse engineering. Bioinformatics 2000, 16: 707–726. 10.1093/bioinformatics/16.8.707View ArticleGoogle Scholar
 Park T, Yi SG, Kang SH, Lee SY, Lee YS, Simon R: Evaluation of normalization methods for microarray data. BMC Bioinformatics 2003, 4: 33. 10.1186/14712105433PubMed CentralView ArticlePubMedGoogle Scholar
 Marshall E: Getting the noise out of gene arrays. Science 2004, 306: 630–631. 10.1126/science.306.5696.630View ArticlePubMedGoogle Scholar
 St. Jude Children's Research Hospital (SJCRH) Database on childhood leukemia[http://www.stjuderesearch.org/data/ALL1/]
 Tsodikov A, Szabo A, Jones D: Adjustments and measures of differential expression for microarray data. Bioinformatics 2002, 18(2):251–260. 10.1093/bioinformatics/18.2.251View ArticlePubMedGoogle Scholar
 Szabo A, Boucher K, Carroll W, Klebanov L, Tsodikov A, Yakovlev A: Variable selection and pattern recognition with gene expression data generated by the microarray technology. Mathematical Biosciences 2002, 176: 71–98. 10.1016/S00255564(01)001031View ArticlePubMedGoogle Scholar
 Bolstad BM, Irizarry RA, Astrand M, Speed TP: A comparison of normalization methods for high density oligonucleotide array data based on variance and bias. Bioinformatics 2003, 19(2):185–193. 10.1093/bioinformatics/19.2.185View ArticleGoogle Scholar
 Simon RM, Korn EL, McShane LM, Radmacher MD, Wright GW, Zhao Y: . In Design and Analysis of DNA Microarray Investigations. Springer, New York; 2003.Google Scholar
 Irizarry RA, Gautier L, Cope LM: An R package for analyses of Affymetrix oligonucleotide arrays. In The Analysis of Gene Expression Data. Edited by: Parmigiani G, Garrett ES, Irizarry RA, Zeger SL. Springer, New York; 2003:102–119.View ArticleGoogle Scholar
Copyright
This article is published under license to BioMed Central Ltd. This is an Open Access article distributed under the terms of the Creative Commons Attribution License (http://creativecommons.org/licenses/by/2.0), which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.