Computational and Mathematical Methods in Medicine

Computational and Mathematical Methods in Medicine / 2012 / Article
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Integrative Approaches in Computational Biomedical Imaging

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Research Article | Open Access

Volume 2012 |Article ID 481923 |

Gonzalo Vegas-Sanchez-Ferrero, Santiago Aja-Fernandez, Cesar Palencia, Marcos Martin-Fernandez, "A Generalized Gamma Mixture Model for Ultrasonic Tissue Characterization", Computational and Mathematical Methods in Medicine, vol. 2012, Article ID 481923, 25 pages, 2012.

A Generalized Gamma Mixture Model for Ultrasonic Tissue Characterization

Academic Editor: Huafeng Liu
Received01 Jun 2012
Revised20 Aug 2012
Accepted31 Aug 2012
Published04 Dec 2012


Several statistical models have been proposed in the literature to describe the behavior of speckles. Among them, the Nakagami distribution has proven to very accurately characterize the speckle behavior in tissues. However, it fails when describing the heavier tails caused by the impulsive response of a speckle. The Generalized Gamma (GG) distribution (which also generalizes the Nakagami distribution) was proposed to overcome these limitations. Despite the advantages of the distribution in terms of goodness of fitting, its main drawback is the lack of a closed-form maximum likelihood (ML) estimates. Thus, the calculation of its parameters becomes difficult and not attractive. In this work, we propose (1) a simple but robust methodology to estimate the ML parameters of GG distributions and (2) a Generalized Gama Mixture Model (GGMM). These mixture models are of great value in ultrasound imaging when the received signal is characterized by a different nature of tissues. We show that a better speckle characterization is achieved when using GG and GGMM rather than other state-of-the-art distributions and mixture models. Results showed the better performance of the GG distribution in characterizing the speckle of blood and myocardial tissue in ultrasonic images.

1. Introduction

Among the noninvasive imaging modalities, probably, the most widespread are the ultrasound imaging. The main reason of its success is that it provides a low-cost way to help diagnosing and can be used for many medical applications. However, ultrasonic (US) images are characterized by the presence of a peculiar granular pattern: the so-called speckle.

This term was adopted from the field of laser optics [1] in the early 1960s due to the similarity of the patterns between laser optics and ultrasonics. Although the nature of the speckle in US images stems from a different phenomenon, there still share some similarities. Both patterns come from the random interference of many coherent wave components reflected from different microscopic elements. In the case of US, the volume, the number of effective scatterers, and the acquisition process contribute to the formation of a speckle [2].

The analysis of backscattered echo from tissues needs a proper description of the ultrasonic signals. For this purpose, and due to the random nature of the speckle, several statistical models have been proposed in the literature. This characterization can be used either for segmentation [3], classification [4] purposes or for filtering the speckle itself [58]. The latter usually considers the speckle as an undesired consequence, since it degrades resolution and adds spatial noise to the image. Thus, filtering is commonly applied as a preprocessing step for further segmentation of regions of interest or to extract relevant measures for physiological analysis.

The statistical description of US signals provide an important information of the backscattered echo from tissues. The parameters of the statistical models allow identifying the features of tissues and provides important descriptors for classification. Some of the filtering algorithms relay on a Bayesian approach where an accurate statistical model becomes necessary. As a consequence, modeling the amplitude statistics of US signals has been a very active area.

Several statistical models have been proposed in the last decades. Probably the most wellknown is the Rayleigh model, which is a one-parameter distribution which describes the so-called fully formed (or developed) speckle. This probabilistic distribution describes the behavior of a speckle when a high number of effective scatterers are present in the resolution cell. However, real images show a deviation from this model, this non-Rayleigh behavior can be due to a small number of scatterers in the resolution cell or when there are some dominant components in the cell. The most commonly accepted distributions that try to model non-Rayleigh distributions are the Rice (fully resolved speckle), (partially formed speckle), and Homodyned (partially resolved speckle).

Although, those models are based on physical assumptions of the backscattering process, some other distributions have proven to provide a good performance on real images. This is the case of Gamma [911] and Nakagami [12] distributions. The first is proposed as a two-parameter distribution that describes the result of interpolated/filtered fully formed speckle [9] and also has shown good results in empirical tests among other distributions [10, 11]. The Nakagami distribution proposed by Shankar for the case US characterization [12] is also a two-parameter distribution which generalizes the Rayleigh distribution. This distribution was adopted from the models proposed to describe the statistics of the returned echo radar.

The capability of the Nakagami distribution to model the backscattering from tissues for fully resolved and fully formed speckle made it become the most commonly accepted model for tissue characterization. However, the tails of the probabilistic density functions of Nakagami, K, Rayleigh, or Gamma do not show the impulsive response of speckle which originate heavier tails. In order to describe this impulsive response, a generalized Nakagami distribution was proposed by Shankar in [13]. This is a three-parameter model which has shown a better behavior than the Nakagami or Rayleigh, an expected result since it is a generalization of the other models. However, the generalized Nakagami distribution does not have closed-form maximum likelihood estimates (MLE) and, thus, it makes their use difficult. Note that, though Shankar in [13] said that the MLE can be obtained, the equations used were based on the results from Stacy and Mihram [14], which were calculated by the methods of moments and they also expressed the difficulties of obtaining an MLE: “Closed expressions for solutions to the maximum likelihood equations are highly unlikely.” It is important to note that the results of [14] were obtained for the estimation of the Generalized Gamma (GG) distribution which is essentially the same as the generalized Nakagami distribution but with another parametrization.

The different nature of tissues is reflected in a different response of the speckle. Hence, a mixture model has shown to be a natural strategy for statistically describing the features of tissues. This approach has been previously used for segmentation purposes in the case of Nakagami mixture models (NMMs) by Destrempes et al. in [3], for classification with Rayleigh Mixture Models by Seabra et al. in [4], and for filtering with a mixture of Gamma and Gaussian Mixture Models in [8, 9]. All these approaches make use of the Expectation-Maximization (EM) [15] algorithm to calculate the parameters that better fit the empirical probability distribution function (PDF). This method is particularly useful when the MLE exists since it maximizes the expected value of the log-likelihood function with respect to the condition of the belonging to each tissue class for a given data.

The EM algorithm cannot be easily applied for the calculation of a Generalized Gamma Mixture Model (GGMM) without an MLE. However, some interesting results have been recently published on the calculation of the MLE of the Generalized Gamma which permit efficient computation of the GGMM.

The aim of this work is to revitalize the use of the Generalized Gamma distribution (also called, Generalized Nakagami Distribution) for tissue characterization. For this purpose, we present two main contributions: first, we propose a simple methodology to calculate the ML estimate which offers robust results comparing to the methods in the literature [14, 16, 17]. Second, two different methods were proposed for the calculation of the GGMM parameters. Both were developed by applying the EM method in the derivation of the proposed ML method. Results when comparing both methods to the GMM and NMM in real images showed the better fitting of the GGMM. The GGMM provides a posterior probability of belonging to each tissue class which can be of help for further filtering, segmenting, or classifying methods.

The rest of the paper is structured as follows. In Section 2.1, we introduce the distributions most commonly used for characterizing speckle of ultrasonic images. There, the GG distribution is motivated as a suitable generalization of the Gamma and Nakagami distributions which fail in describing the impulsive response of speckle. Then, in Section 2.2, we analyze the methods proposed in the literature for estimating the parameters of the GG distribution and a simple but robust method is proposed (Section 2.2.4). One of the advantages of this method is that it can be easily extended to estimate the parameters of a GGMM by means of the EM algorithm. Section 2.3 is devoted to the extension of the ML method to obtain the parameters of the GGMM where two algorithms are proposed. The performance of the ML estimate derived in Section 2.2.4 is compared to other state-of-the-art methods in Section 3.1 for synthetic data and for real cases in Section 3.2. The performance of the GGMM is analyzed in Section 3.3, where the GGMM is compared to NMM and GMM. Finally, we propose some applications for the GGMM in Section 3.4. In Section 4, we conclude.

2. Materials and Methods

2.1. Statistical Models for Describing the Nature of Speckle

The formation of US images begins with the emission of a pulse packet which travels through the tissue. The backscattering produced by the scatterers in the resolution cell contribute to the change of the pulse shape according to the characteristics of the media, that is, the number of scatterers as well as their size [4, 9, 12].

The contribution of the backscattered echo, , can be treated as a random walk due to the random location of the scatterers in the resolution cell [12]: where is the mean frequency of excitation and is the number of effective scatterers in the resolution cell. The phases, , are usually modeled as uniformly distributed in and the amplitude is usually considered to be Normal distributed.

The fully formed speckle model assumes a high number of scatterers, so the Central Limit Theorem applies and the backscattered echo can be expressed as where and are zero mean identically distributed Gaussian distributions.

Then, the envelop of the backscattered signal echo, is Rayleigh distributed [1, 18]: where is the Heaviside step function defined as

Under the assumption of a high number of effective scatterers but with the presence of resolvable structures in the resolution cell (specular component, ), and become nonzero Gaussian distributions. The envelop does no longer follow a Rayleigh distribution but a Rician one [18]: where is the modified Bessel function of first kind.

When the number of scatterers decreases and the Central Limit Theorem cannot be applied, more complicated distributions are proposed to model the distribution of the envelope. Concretely, the distribution models the case when the number of scatterers is a random variable itself, which is modeled as a Poisson whose local mean is Gamma distributed, this is equivalent to consider as gamma distributed [2]: so, the PDF of is where is the modified Bessel function of the second kind.

A generalization of the previous models appears when a specular component is considered and the number of scatterers, , follows a negative binomial distribution. This is the case of the homodyned-K distribution [19]: This PDF has no closed expression and this limits its use.

On a completely different approach, Shankar in [12] proposed a Nakagami distribution as a “simpler universal model for tissue characterization.” Unlike the previously reviewed models, the Nakagami is not based on physical arguments or on a bottom-up modeling of the scattering process. However, it has empirically shown a better performance than the Rayleigh and Rice distributions.

The Nakagami PDF is as follows:

This distribution offers good properties to describe the backscattered echo: the Rayleigh distribution is a particular case of the Nakagami () and, additionally, when is similar to the Rice distribution. However, this distribution has some limitations. The Nakagami model cannot fit the heavier tails of the empirical PDFs due to the impulsive nature of scatterers [13].

In order to describe the impulsive response of scatterers, Shankar proposed in [13] a generalized Nakagami distribution which is essentially the same as a Generalized Gamma distribution [14]. However, this distribution presents some difficulties in the estimation of its parameters, since there are no closed equations for the ML estimates.

In the next section, we describe some methods that have been used in the literature with special attention to methods that provide an ML estimate of the GG parameters. Additionally, we propose a simple method to calculate the ML estimates of the parameters. The results obtained in the derivation of this ML method provide the foundations for the development of the Generalized Gamma Mixture Model, which is the main contribution of this work.

2.2. Estimation of Parameters of the Generalized Gamma
2.2.1. Moments Method

This method was proposed by Stacy in [14]. For the derivation of the method, the following parametrization was adopted: where the parameters are all positive.

This is the definition of the GG distribution hereafter. For a given , all moments exist.

Now, let be the random variable (RV) defined as

For this RV, the central moments, , of rth order are

Additionally, it is easy to show that, given a RV, , which follows a GG distribution (), the following properties hold:

So, a new RV can be defined as where follows a Gamma distribution of parameter . Hence, the log-transformed distribution of the Gamma RV is the following: where .

The moment generating function of is easily calculated as . Where is the expectation operator. So, the th moment of is the following: where is the polygamma function defined as

Finally, the three first central moments are defined as:

These equations can be used to estimate the parameters of the ; , by approximating the moments by means of the sample moments: where , with the set of samples of ; is the sample variance of , and its sample skewness.

The estimates are derived by means of calculating the value from the last equation of (18). So, a numerical calculation needs to be performed. In the original article [14], Stacy and Mihran provided a graph representing for a range .

This method, though provides a quite straight-forward calculation of the parameters, can provide estimates which are outside the parameter space. Yet, it is highly sensitive to the number of samples.

2.2.2. Heuristic Approaches

In order to avoid the problems associated to the moments method, some heuristic methods have been proposed in the literature. As examples, Gomes et al. [16] proposed an iterative method which evaluates the best performance of the goodness-of-fit test for a fixed (see the parametrization of (10)). The parameters of the transformed samples , which are Gamma distributed, were calculated by the moments method. At the end of the loop, the set of parameters with least value is chosen.

This method presents some shortcomings. First, the parameters of the Gamma distributed data were calculated by the moments method, so the problems associated to the moments method are not circumvent. However, even if a good estimate is calculated, the goodness-of-fit test depends on the calculation of the estimated PDF which strongly depends on the number of bins considered and the assumption of a sample with sufficient large size.

Other heuristic method is the one presented by Wingo in [20]. This method, based on the one proposed by Hager and Bain [21], tries to solve the maximum likelihood equations for the GG distribution. The log-likelihood, , of a RV for the parametrization presented in (10) is where is the set of samples.

Now, calculating the derivatives with respect to the parameters and setting it equal to zero, one can obtain the ML equations: where .

This system of equations can be reduced to a single nonlinear equation with as the single unknown: where So, the problem is reduced to calculate from (21). Some authors reported the difficulty of solving this equation with the conventional numerical methods such as Newton-Raphson [21] and conclude that the MLE may not exist.

In [20], the author faced the problem by analyzing the effect of inappropriate zero finding algorithms. So, an heuristic method for isolating roots of a general scalar nonlinear equation was proposed. This method makes use of the root-isolation technique proposed in [22], which uses only function values to isolate the roots in a compact interval of the real line.

Though this method can provide an ML estimate of the parameter by solving (21), it has to heuristically divide the intervals where is searched and calculate whether a root is in it or not by means of the mean value and variance of the function in each of the intervals, so many evaluations of the function are required.

2.2.3. ML Approach

A very interesting analysis was recently published by Noufaily and Jones in [17], where an iterative approach is proposed to solve the likelihood equations, (20), in a way that the individual equations are uniquely solvable. This result provides a very promising technique for calculating the MLE parameters of the GG.

In that work, the log-likelihood equations were calculated following the re-parametrization proposed in [23]. Concretely, for a RV which is distributed, , the new RV is calculated, whose PDF is the following: where , , and .

So, in the end, the following equations have to be solved: where and .

The important result of [17] is the demonstration that both (25) and (26) are well behaved with unique solutions in and , respectively. So, an iterative method can be developed to calculate by (26) from an initial guest of the parameters and then by solving (25). Finally, is calculated by replacing the previous estimates in (24). These estimates can be used to calculate a new to compute the new log-likelihood function. By repeating these steps until a desired accuracy, the estimates are achieved [23].

This method provides a fundamental result about the behavior of the log-likelihood equations, and guarantees their solution. However, the method does not provide any proof concerning its convergence or the uniqueness of the ML. Yet, this method needs to solve two nonlinear equations by numerical techniques, whereas the method proposed by Wingo in [20], previously described, only needs to solve a linear equation.

2.2.4. The Proposed Approach

We propose a simple but efficient method to calculate the ML estimates of the GG distribution. The main advantage of the method is that it can be easily implemented and has the same properties of the method of [17], that is, the equation to solve are well behaved with unique solution. Additionally, the method just needs the calculation of just one non-linear equation and, thus, the computing time is considerably reduced.

The method consists in transforming the RV, by the following transformation where is a positive real number. So, the new PDF of is the following: Note that this PDF follows a Gamma PDF when . Hence, a reasonable way to find the value is to find the value of that maximizes the likelihood of the GG distribution and also maximizes the Gamma distributed RV .

In order to see if this method provides a proper solution, we first demonstrate that the ML estimate of the parameters of the new random variable also maximizes the likelihood of the GG distribution when .

First, we calculate the ML estimates of the parameters of (27) for , whose log-likelihood is the following:

The maximum with respect to the parameter is easily calculated by taking the derivative with respect to the and setting it equal to zero: Finally, (28) can be maximized with respect to by introducing the value of :

Now, by introducing in the log-likelihood function of the GG distribution, (19): Now, by maximizing with respect to , we obtain the following equation: and finally, reordering terms, we obtain the same equation for which is also a solution.

This result guarantees that there exists always a solution for the ML estimate of the GG distribution (, , ) and the parameters and are those obtained for the ML estimate for the transformed RV . Hence, there is always a solution for the MLE for a GG.

Additionally, since the MLE of a Gamma distribution always exist for whatever positive values ((30) is well behaved), the problem is reduced to finding the value that maximizes among the ones that maximize .

The search method for was implemented by the Nelder-Mead method [24] while the Brent’s algorithm was applied for calculating [25].

This method does not demonstrate the uniqueness of as did not any of the methods in the literature. However, in our experience, we agree with Noufaily and Jones [17] that the global maximum of the appears to be distant to any other local maximum.

The main advantage of the method here proposed is that it is easy to implement and only one non linear equation has to be solved, whereas the method of [17] needs to solve two non-linear equations in each iteration and [20] method needs several calculations of non-linear equations for each interval considered for the isolation root technique.

2.3. Generalized Gamma Mixture Model

An additional advantage of the proposed method for the calculation of the MLE parameters for the GG distribution is that it can be easily adapted for the calculation of the parameters of GG Mixture Models (GGMM).

There were some attempts in the literature to obtain the parameters of a GGMM. Concretely, in [26], they calculated the GGMM by means of the Nelder-Mead and Gradient descent methods for maximizing the log-likelihood. However, that method is strongly sensitive to the number of mixtures since it is just a direct optimization of the log-likelihood score equations of the mixture model.

In this section, we derive the GGMM by applying the Expectation-Maximization methodology [15] and combining them with the method used to calculate the MLE of the GG distribution.

Let , be a set of samples. These samples are considered to be independent and identically distributed (IID) RVs. Now, the GGMM considers that these samples result from the contributions of distributions: where is a vector of the parameters of the GGMM and are the parameters of the PDF (in our case the parameters of the GG, represented as , , ).

The joint distribution of IID samples is given by

The EM is applied here to maximize the log-likelihood function when some hidden discrete random variables are introduced into the model. These RVs take values in and indicate the class for which each sample belongs.

Now, defining as an estimate of the parameters of the mixture in the th iteration, the expectation step is performed by calculating the expected value of the log-likelihood :

In the maximization step, the new estimate is obtained by maximizing the expectation of the log-likelihood function . These steps are iterated until a stop criterion such as for some preestablished tolerance (Tol) is reached.

The application of the EM algorithm for estimating the parameters of mixture models has been applied for several distributions, see, for example, [15, 27]. However, to the best of our knowledge, this is the first time a mixture model is presented for GG distributions.

In order to derive the estimates of the parameters in each iteration, we first define the joint distribution of IID samples and the hidden random variables, as where .

Now, the log-likelihood function can be defined in the following way:

The expectation of the log-likelihood function with respect to the hidden RVs when data and the previous estimate are known as: where is the probability of to belong to the class .

The probability can be calculated by applying the Bayes theorem as

Note that (37) is composed of two terms, so the maximization step can be done to each term independently. For the term depending on the some constraints have to be considered since they must hold . An optimization via Lagrange’s multipliers can be done in a straightforward way and they guarantee a necessary condition for optimality in this problem. The new Lagrange function with as the Lagrange multiplier is the following: where we introduced to simplify notation.

Now, calculating the derivative with respect to each and setting it equal to , the following expression is derived:

By summing both terms of the equation over , we obtain and the estimates for the parameters that maximize the Lagrange function (and the likelihood function) are

For the calculation of the maximum of (37) with respect to , we first calculate the derivative with respect to : where the log-likelihood of is the one described in (19) for one sample :

The result is

Now, plugging (45) into (37) and deriving with respect to and setting it equal to It results in the following equality:

Note that (47) is essentially the same as (30), which is well behaved and always has a unique solution. Thus, this nonlinear equation can be solved by numerical methods in the same way as was performed in the MLE of the GG parameters. In our case, we also used the Brent’s algorithm [25].

The interval where the Brent’s algorithm is performed can be derived by means of the following property:

So, the desired value of in the interval where

This property can be found in [28] and was also used in [17] for the calculation of the ML estimates of the GG.

Now, the problem can be stated in the same way as was done for the ML estimate proposed in Section 2.2.4. We are interested in the parameter which maximizes the likelihood for the component . So, for each , (45) and (47) provide the estimate of and , respectively. By applying the Nelder-Mead algorithm to maximize the log-likelihood function for each component , as was done for the ML estimates in Section 2.2.4, one can obtain the desired ML estimates. We will refer to this method as the GGMM1 method.

It is important to note that the parameter estimates can be also solved by extending the ML method of [17]. For this purpose, the parametrization proposed by Lawless [23] can be applied to the mixture model as was explained in Section 2.2.3.

The log-likelihood equations to be solved are completely equivalent to (45) and (47) due to the invariance of the ML estimates to the transformation . However, Lawless’ parametrization allows us to extend the results of [17] to the case of GGMM. For the sake of clarity, we rewrite the parametrization: With this parametrization, (45) becomes where

So, in the case of the parameter which maximizes the log-likelihood of : It results in

This equation is well behaved and all the theoretical demonstrations obtained in [17] still hold: it is monotone decreasing and, when , the function takes the value

As a conclusion, (55) has always a positive solution for any and . Additionally, due to the invariance of the ML estimates for the transformation , there always exist a for any and .

The solution is in the interval So, the value can be calculated by any numerical method. We used here also the Brent’s algorithm.

So, finally, from an initial guess of one can calculate from (47) and then use it to calculate the estimate of from (55), in an iterative way until a desired tolerance is reached.

This methodology generalizes the proposed method of [17] for the case of GGMM and we will refer to it as the GGMM2 method.

2.4. Implementation Generalized Gamma Mixture Model

In this section, we detail the implementation of both of the proposed methods for the GGMM.

In Algorithm 1, the Nelder-Mead method [24] was used for the calculation of and the Brent’s algorithm [25] for in the interval given in (49).

Number of components
Initial guess of parameters for
each component
maxIter Maximum number of iterations
Tol Tolerance
Iterations counter
while err > Tol and < maxIter do
for     do
end for
 err Evaluate the relative tolerance
end while

In the case of Algorithm 2, the Brent’s algorithm [25] was used for calculating and in the intervals of (57) and (49), respectively.

Transformed samples
Number of components
Initial guess of parameters for
each component
maxIter Maximum number of iterations
Tol Tolerance
Iterations counter
maxIterML Maximum number of iterations for the ML algorithm
TolML Tolerance for the ML
while err > Tol and < maxIter do
for     do
   Iterations counter for ML
   Calculate the log-likelihood for the th component
  while errML > TolML and < maxIterML do
    Calculate the log-likelihood for the th component
  end while
end for
 err Evaluate the relative tolerance
end while

The computational complexity of the previous GGMM methods when compared to the calculation of a simple GG depends on the number of components, , assumed by the model. In each iteration of the EM algorithm, the expected parameters of each component have to be calculated. So, if the time consumed to estimate a GG is , the calculation of the expected parameters of the mixture is .

When other mixture models such as RMM, NMM, and GMM are considered, the complexity of the EM method is similar to the GGMM. Note that both the GMM and the NMM need to solve a non-linear equation similar to (47) so the consumed time of the solution is the same. The additional cost of calculating the GGMM parameters is due to the calculation of the estimate of the parameter . If we define the time to solve (47) and (45) as , and as the time consumed for solving , the computational time for a simple GG () and a GGMM of components () would be

The estimated times in a Matlab (R2011a) implementation running in an ASUS G53SW laptop (Intel Core i7 2630QM Processor, 2.2 GHz, 8 GB RAM) were  ms and  s.

3. Results and Discussion

3.1. Performance of the ML Method

In this section, we show the performance of the proposed methods for calculating the parameters of a GG distribution. For this purpose, we performed synthetic experiments and tested the methods presented in Section 2.2. Concretely, we tested the method of Stacy and Mihram [14], Gomes et al. [16], Noufaily and Jones [17], and our proposed method of Section 2.2.4. We will refer to them as Stacy, Gomes, Noufaily, and proposed methods, respectively.

The synthetic data was calculated in the same way as was done in [17]: a set of gamma-distributed random samples are generated by means of the method proposed in [29] and the GG-distributed data are obtained by taking the th power of the samples. The parameters of the GG distribution were also calculated from sets of parameters in a reasonable dynamic range. The scale parameter was set to in all the experiments since this parameter just affects to the scale of the data. Both, the parameter and the parameter were obtained from random samples of a uniform RV in the interval .

We choose this interval since lower values than make the distribution to take values that tend to infinity as get closer to . This is an unrealistic situation when real images are considered. Additionally, when takes lower values, the tail becomes heavier and the shape of the distribution also becomes unrealistic. These effects are shown in Figure 1, where some examples of the PDFs of the GG distribution are depicted.

The number of iterations for the proposed method and for the method of [17] was set to and the tolerance function to . The number of bins where the -test was performed in the method of [16] was and the number of samples per experiment was . The comparisons of the methods were performed by comparing the goodness-to-fit of each distribution by means of two different measures: Kullback-Leibler divergence (KL) and Kolmogorov-Smirnov (KS) statistic. The former is a nonsymmetric measure of the difference between two probability distributions defined as where is the empirical PDF estimate and is the theoretical distribution (the GG distribution). For the empirical estimate of the PDF, the number of bins of the histogram was set to 150.

The Kolmogorov-Smirnov statistic is the uniform norm of the cumulative distribution function (CDF), defined as where is the empirical CDF of data and the theoretical CDF. The KS measure was chosen since it does not depend on the PDF estimate and can be calculated with a few number of samples. Additionally, the Glivenko-Cantelli theorem states that if the samples are drawn from distribution FX, then converges to 0 almost surely [30].

In Figure 2, the results for both measures are depicted. It is clear that the moments method of Stacy gives poorer results than the other methods for both measures. This result was expected since the moments method depends on moments of third-order, so the variance of the estimates becomes higher. The rest of the methods performed well for both measures. In the case of the , they fit practically the same while, in the case of , there are some better results for the method of Noufaily and the proposed one. This is the effect of the approximation of the PDF for the test performed by the method of Gomes: it calculates the best set of parameters for an approximation of the empirical PDF which depends on the number of bins and the number of samples of the dataset. So, as the number of samples is reduced or the number of bins is reduced, the estimate becomes worse.

In order to see the effect of this, we also show in Figure 3 the relative error of the estimates for all the methods (the relative error of an estimate is calculated as , while the absolute error is ). In the figure, the whiskers show the dynamic range of the data which is not considered an outlier. So, though the method of Gomes provides good fitting, the variance of the estimates is higher than the method of Noufaily and the proposed one. At first sight, the results of Figure 3 demonstrate the better performance of the proposed method in terms of variance of the ML estimates with no appreciable bias in the estimates.

An example of the fitting performance of the methods is shown in Figure 4 where the PDFs obtained with the methods are depicted as well as the absolute error and the relative error of the PDFs.

Following, we analyze the dependence of the estimates with the number of samples. The same experiment is repeated considering 500 samples. The results of both goodness-to-fit measures are shown in Figure 5, and the relative errors of the estimates are depicted in Figure 6. The performance for the measure is similar for all the methods. However, note that the value is considerably higher than the obtained for the case of samples, this effect is caused by the difficulties of estimating the PDF with so few samples. Since the Gomes algorithm is based on the test, it is expected that its performance decreases and the variance of the parameter estimates increases. In the case of the measure, the performance of all methods is comparatively equal to the case of samples but a higher variability is observed in the Gomes method due to the sensitivity to the number of samples.

The better performance of Noufaily and the proposed methods are seen in Figure 6 where the variability of the Noufaily method did not increase dramatically as the Gomes method did. The proposed method also presented a very low variance of the parameter estimates with no appreciable bias. In the light of these results, we can conclude that the proposed method is robust with respect to the number of samples and it does not introduce any appreciable bias in the parameter estimates. The goodness-of-fit performance of both the Noufaily and the proposed method are similar, though the estimates are more accurate with the proposed method. This can be due to the better convergence of the Nelder-Mead method than the algorithm of the Noufaily method.

3.2. Tissue Characterization in Real US Images

In this section, we test the performance of the GG distribution for characterizing tissues of real images. For this purpose, we used a set of 518 real US images (, 8 bits) obtained from 3 human subjects by means of a clinical machine GE Vivid 7 echographic system (GE Vingmed Ultrasound AS, Horten, Norway). The images were acquired before the Cartesian rearrangement. The image collection was supervised by specialists Marta Sitges and Etelvino Silva (Hospital Clinic IDIBAPS Universitat de Barcelona, Spain). The subjects were volunteers for a study of the reconstruction process of ultrasonic images. The acquisition was done in the Hospital Clinic of Barcelona with its approval. The images were provided by Nicolas Duchateau (CISTIB-Universitat Pompeu Fabra, Ciber-BBN, Barcelona, Spain) and Bart Bijnens (Instituco Catalana de Recerca i Estudis Avan cats (ICREA), Spain).

In Figure 7, an example of a real US images is shown with its Cartesian rearrangement. The red contour is the segmented areas of blood which are considered in the study, while the green contour is the segmented areas of tissue. The intersection of both regions was rejected in the study.

Additionally, the histogram of the image was depicted for the blood region as well as the fitted distributions most commonly used to characterize tissue. From the whole data set, a total number of regions were segmented for myocardial tissue while were segmented as blood. The sizes of regions vary depending on the tissue. However, it is high enough to provide a good estimate of the parameters. For instance, the segmented region of Figure 7 has samples for blood and for tissue.

In the case of Figure 7, the lower value of the histogram shown is since the intensity values in the blood area were in the interval . The number of bins used for the representation of the histogram was set to equally spaced in that interval.

The performance of the GG distributions was tested by estimating the PDFs for both tissue classes (myocardial tissue and blood) for the following distributions: Exponential, Rayleigh, Weibull, Normal, Nakagami, Gamma, and GG. The PDFs were compared by means of both the and the measures. The results of the comparison are depicted in Figure 8 where the better performance of the Gamma, Nakagami, and GG becomes clear. In order to see whether these measures are statistically significant, we carried out a Welch -test for the Gamma, Nakagami, and GG distributions for the measures. This test was chosen since no equal variance should be assumed and the since it does not depend on the empirical PDF estimate but just on the samples. The assumed hypothesis is that “both distributions have the same mean,” indicates that the null hypothesis can be rejected at a of significance level.

The results are shown in Tables 1 and 2. Note that all the null hypothesis were rejected but just one: myocardial tissue. In that case, the difference of the mean value of the Gamma and the GG is not statistically significant. The mean values of the are represented in Table 3 where the lower mean value of the GG for both tissues can be appreciated. The results of the -test of Tables 1 and 2, and the lower mean values of the evidence the better performance of the GG than the rest of the distributions, with the exception of the myocardial tissue, where a Gamma distribution offers the same performance.

Blood value Hypothesis

Nakagami versus Gamma <10−15
Gamma versus GG
Nakagami versus GG <10−15

Myocardial tissue value Hypothesis

Nakagami versus Gamma
Gamma versus GG
Nakagami versus GG

Nakagami Gamma Generalized Gamma


3.3. Performance of the GGMM Methods

In this section, we test the performance of the proposed GGMM methods in three different scenarios. First, we test the necessity of using more than a simple GG for describing tissues with an increasing echolucent response of the effective scatterers. The case of a variation of the number of effective scatterers is also considered. This behavior can be found in structures with an increasing deterministic response that changes the speckle nature from fully formed speckle to fully resolved speckle. The variation of the number of effective scatterers can be found in structures which change their scattering cross-section.

In order to simulate B-mode US images, we followed the same methodology proposed in [9]. This method scans an image and records the data in a matrix which is corrupted by means of the speckle formation model of (1) where the tissue is modeled as a collection of scatters of size comparable to the wavelength. The speckle pattern is obtained by means of a random walk which does not assume any statistical distribution in order to avoid any bias of the results. The Cartesian arrangement is obtained by means of linear interpolation of the corrupted samples.

As a first example, we simulate an increasing echolucent tissue which varies its intensity from 0 to 255 from left to right. The sampling process and the resulting B-mode image are shown in Figures 9(a)-9(b). The number of samples were set to 50 angular samples and 100 radial samples, represented as red points. The amplitude of each scatterer is defined as a Normal distributed RV with zero mean and . Note that, along with the variation of intensities from left to right, a specular component of the speckle will appear. The number of scatterers was set 20 in order to simulate fully formed speckle in regions with low echolucent response and fully resolved speckle in regions with high echolucent response. The resulting B-mode image is represented in Figure 9(b).